research studies evaluating how effective individual therapies are at

• Research article
• Open access
• Published: 11 January 2013

The effectiveness of individual interpersonal psychotherapy as a treatment for major depressive disorder in adult outpatients: a systematic review

• Madelon L J M van Hees 1 ,
• Thomas Rotter 1 , 2 ,
• Tim Ellermann 3 &
• Silvia M A A Evers 1 , 4  

BMC Psychiatry volume  13 , Article number:  22 ( 2013 ) Cite this article

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This systematic review describes a comparison between several standard treatments for major depressive disorder (MDD) in adult outpatients, with a focus on interpersonal psychotherapy (IPT).

Systematic searches of PubMed and PsycINFO studies between January 1970 and August 2012 were performed to identify (C-)RCTs, in which MDD was a primary diagnosis in adult outpatients receiving individual IPT as a monotherapy compared to other forms of psychotherapy and/or pharmacotherapy.

1233 patients were included in eight eligible studies, out of which 854 completed treatment in outpatient facilities. IPT combined with nefazodone improved depressive symptoms significantly better than sole nefazodone, while undefined pharmacotherapy combined with clinical management improved symptoms better than sole IPT. IPT or imipramine hydrochloride with clinical management showed a better outcome than placebo with clinical management. Depressive symptoms were reduced more in CBASP (cognitive behavioral analysis system of psychotherapy) patients in comparison with IPT patients, while IPT reduced symptoms better than usual care and wait list condition.

Conclusions

The differences between treatment effects are very small and often they are not significant. Psychotherapeutic treatments such as IPT and CBT, and/or pharmacotherapy are recommended as first-line treatments for depressed adult outpatients, without favoring one of them, although the individual preferences of patients should be taken into consideration in choosing a treatment.

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Major depressive disorder (MDD) is a mental disorder characterized by a depressed mood, diminished interest or pleasure, sleeping problems and tiredness, and negative thoughts [ 1 ]. The mean one-year-prevalence of depression in European inhabitants between 18 and 65 years old is 6.9% [ 2 ], and 16.2-16.6% of US adults develop a major depressive disorder [ 3 , 4 ]. Furthermore, depression causes a high burden worldwide, taking fourth place in a ranking of leading contributors to the burden of diseases in 2000. In 2020, it is estimated that depression will take second place in the ranking for all ages and sexes [ 5 ]. Moreover, depression is the leading cause of years of life lived with disability, in all ages and sexes, accounting for 11.9% of all disability [ 6 ]. Since it appears that persons suffering from mental disorders make more use of health care services [ 7 ], the increasing prevalence of depression leads to an increase in health care costs.

Research [ 8 ] and Dutch guidelines [ 9 ] suggest treating depression with psychotherapy and/or medication. Psychotherapy follows several kinds of methodologies. For depression, Cognitive (Behavior) Therapy (CBT) and Interpersonal Psychotherapy (IPT) are often applied. CBT originates from behavior therapy and cognitive therapy, and combines elements of both therapies [ 10 – 12 ]. IPT was originally developed for treating acute depression by improving the interpersonal functioning with important others [ 13 – 17 ]. This study will focus on the effectiveness and efficacy of IPT, since CBT has been subject of many studies up until now, while IPT has only recently become a subject of interest.

As a monotherapy for adults, individual IPT appears to be an effective treatment for depression [ 18 – 20 ], and several reviews [ 21 – 25 ], and meta-analyses [ 26 – 33 ] have been performed on the effectiveness of all kinds of methodologies of psychotherapy. Nevertheless, psychotherapy is a broad concept, and reviews and meta-analyses have often focused on different combinations of psychotherapy for treating depression without comparing one specific sole treatment to another [ 21 , 25 – 30 , 32 , 34 ]. Furthermore, although sole individual IPT appears to be effective, few reviews focus on sole individual IPT in adults with MDD as a primary diagnosis. Often, dissimilar study populations are compared with each other, for example adult, adolescent, and elderly patients in one study [ 23 , 25 – 30 , 33 – 35 ]. Furthermore, several more types of depression exist: dysthymic disorder or depression with medical conditions, for example, but this review will focus only on MDD. Chronic MDD and postpartum depression (PPD) will be included in this systematic review, for the following reasons. First of all, treatment for patients with chronic and non-chronic depression is equal in terms of content and structure. Therefore, the treatments of these patient groups are comparable. Secondly, the symptoms of both kinds of depression are comparable in terms of severity and content, which makes the patients comparable. Furthermore, women with PPD experience the same kind of symptoms as patients with MDD.

Since comorbidity is very common in patients suffering from depression, and this possibly increases the severity of the depression [ 36 – 44 ], this review will focus on MDD as a primary diagnosis with possible comorbidity.

Other factors influencing the results of previously executed systematic reviews include different age groups, in which form the provided IPT is administered, distinct settings, and the time periods during which the studies were executed. IPT is often adjusted for applicability to elderly [ 45 ] or adolescent [ 46 ] depressed patients, or in the form of group IPT [ 47 ]. Therefore, these kinds of treatments may be hard to compare with each other. That is why this review focuses on individual IPT. From here on, when IPT is described in the review, individual IPT is meant, unless described otherwise. Furthermore, the setting in which treatment takes place suggests the depression’s level of severity. It is assumed that inpatients have a more severe depression, which is harder to treat. In addition, inpatient care is often multidisciplinary, which makes it difficult to examine the effects of separate therapies. Research has been conducted on IPT since the 70s, which is why the date limit for this review is set on 1970. This review will give an overview of studies published between January 1970 and August 2012, with a focus on sole IPT administered to adults. Since some therapies have an effect relatively quickly, we did not apply a minimum for duration of a therapy.

With all of the above in mind, the aim of this study is to give an overview of recent literature describing the effectiveness and efficacy of sole individual IPT in comparison with standardized forms of treatment for treating patients with MDD as a primary diagnosis. The following research question has been formulated: Is individual interpersonal psychotherapy more preferable in comparison with other standardized forms of treatment for treating adult outpatients with a primary diagnosis of major depressive disorder?

In order to answer this question, a systematic review will be performed on RCTs and C-RCTs comparing the effectiveness (the outcome of a new treatment compared to other kinds of treatment(s), usually in a clinical setting) or efficacy (the outcome of treatments in homogeneous patient groups, usually in an experimental setting) [ 48 ] of individual sole IPT with other standardized forms of treatment, for treating adult outpatients with MDD as a primary diagnosis.

This paragraph will outline which steps were taken in order to perform this systematic review. An overview of the methods used for data collection, study selection, and data analysis will be provided.

Data sources

RCTs about IPT for depression were collected by searching PubMed and PsycINFO for studies published between January 1970 and August 2012. The following medical subject heading (MeSH) categories and keywords were used: depression, postpartum depression, major depressive disorder, dysthymic disorder, interpersonal psychotherapy, treatment outcome, clinical trials. The exact search terms and MeSH headings can be found in the additional files (Additional file 1 – Search strategy). All titles and abstracts were screened, and only studies which met the review inclusion criteria (see next paragraph and Table 1 ) were selected for further review. Citation tracking and snowballing techniques added studies to the second screening phase, in which selected studies were screened for eligibility using a predefined checklist (see Data analysis) (Additional file 2 – Checklist).

Study selection

Only studies with sufficient methodological quality meeting the inclusion criteria were selected for this review. The criteria for selection will be described shortly. An overview of the inclusion and exclusion criteria is provided in Table 1 .

Studies were included if they were randomized or cluster-randomized evaluations (RCTs or C-RCTs) published in English after January 1 st , 1970, and took place in western jurisdictions, to ensure high internal validity. These studies had to focus on MDD (non-chronic or chronic) as a primary diagnosis in adults (18–65 years old). The diagnosis must have been reached using a formal classification system, such as the Diagnostic and Statistical Manual of Mental Disorders (DSM) [ 1 ], the International Classification of Diseases (ICD) [ 49 ], or the Research Diagnostic Criteria [ 50 ]. Bipolar disorders as primary diagnoses were excluded, as well as cases where the patients were elderly people or adolescents, or in cases in which physical conditions might contribute to the (severity of) depressive symptoms. The proposed intervention must have been individual sole IPT, in comparison with other psychotherapies, pharmacotherapy, or combined treatment. Group IPT and other kinds of treatments were excluded. Studies executed in ambulant care or primary care were included, whereas inpatient care patients were excluded.

By making the inclusion criteria very strict, a more homogeneous group, with a narrower scope, was created, which made it possible to focus on clinical applicability of the treatments for these kinds of patients.

Data analysis

Before the data were analyzed for this review, the methodological quality of the studies included after screening has been assessed, using a predefined checklist (Additional file 2 – Checklist). This checklist was composed of Delphi-list questions [ 51 ] and questions assessing the risk of bias in effect evaluation studies [ 52 ]. General questions were composed for collecting relevant information about the study, after which the resulting information was entered in a Microsoft Excel table for a clear overview. This overview was used to create a table of evidence of the extracted study data, and to summarize the most important findings. MH performed the analysis and consulted TR in case of doubt. In this case, the analyses were double checked and consensus was reached.

The literature search resulted in 3981 studies, of which 3911 were excluded from further review for several reasons, documented below. Figure 1 shows the flow diagram of included and excluded studies. Studies were excluded when they did not meet the inclusion criteria, based on the title and abstract: i.e. they did not focus on MDD as a primary diagnosis, on individual sole IPT, or the target group was anything other than adults. Another 62 were excluded after reading the full text, leaving 8 articles eligible for this review.

Flow diagram of included and excluded articles; reasons in Additional file 3 .

These 62 full-text articles were excluded for the following reasons: being reviews or meta-analyses [ 21 – 23 , 25 – 30 , 32 , 34 , 53 – 60 ], being a protocol for a study [ 61 ], being a study based on earlier/other studies [ 43 , 45 , 62 – 76 ], there was no comparison in the study [ 77 ], MDD was not the primary diagnosis [ 78 – 82 ], the study had the wrong aim for this review [ 83 – 86 ], there was no research data [ 87 – 91 ], or one of the interventions was not IPT as described in the eligibility criteria [ 35 , 92 – 100 ]. See Additional file 3 – List of excluded studies for a detailed description of the reasons for exclusion.

Description of the studies

The main characteristics of the RCT studies included are summarized in Table 2 . One study was carried out in the Netherlands [ 101 ], one in New Zealand [ 102 ], one in Canada [ 103 ], one in the UK [ 104 ], one in Germany [ 105 ], and three in the USA [ 106 – 108 ]. All studies clearly described eligibility criteria and success-of-treatment point. All but two [ 103 , 104 ] included an intention-to-treat analysis. Seven studies reported comparable sociodemographic and psychiatric variables at baseline. One [ 103 ] did not report these variables.

A total of 1233 patients were included in the review, of whom 854 completed treatment in outpatient facilities. Of the patients included, 392 received IPT, 14 received CBASP (Cognitive Behavioral Analysis System of Psychotherapy), 160 received CBT, 153 received pharmacotherapy (nefazodone, nortriptyline hydrochloride, or venlafaxine hydrochloride), 67 received pharmacotherapy plus clinical management, 49 received IPT and nefazodone, 47 received IPT and a placebo, 34 received a placebo plus clinical management, 92 received usual care consisting of communication with a physician for appropriate treatment, and 51 were put on a wait list. The mean age in seven studies [ 101 , 102 , 104 – 106 ] ranged from 29.4 to 40.2 years old, and the percentage of female patients varied from 55% to 83%, except for one study, in which only females participated [ 108 ]. One study did not report these data [ 103 ]. All patients were diagnosed with non-psychotic MDD as a primary diagnosis according to the DSM-III-R [ 109 ], DSM-IV [ 110 ], or the Research Diagnostic Criteria [ 50 ].

IPT in all studies was based on a standardized manual [ 14 , 17 ], as was CBASP [ 111 ] and CBT [ 12 , 112 , 113 ]. The number of IPT and CBT sessions varied from 8 to 24 in a 12- or 16-week period, and most of the sessions were held weekly. Physicians administering nefazodone or nortriptyline were instructed to follow a manual. Patients receiving nefazodone started at 100 mg capsules per day, and doses were gradually increased to a minimum of 400 mg, with a maximum of 600 mg [ 101 ]. Patients receiving nortriptyline started at 25 mg per day, aiming for blood levels of 190–570 nmol/liter [ 106 ]. Patients receiving imipramine hydrochloride had a dosage between 150 and 185 mg. Pharmacotherapists administering venlafaxine followed an evidence-based protocol of 37.5 mg twice-daily doses [ 104 ]. Pharmacotherapy plus clinical management was administered by a psychiatrist who followed the client for the duration of the protocol associated with the antidepressant medication [ 103 ], or as long as the clinical management would be administered [ 107 ].

Risk of bias

Risk of bias was measured and summarized (see Figure 2 ) according to the standards of the Cochrane Collaboration [ 52 ]. Although this was not always described exhaustively, all studies used randomization and seemed to present complete outcome data. Therefore, all included studies had a low risk of selection bias and attrition bias. Nevertheless, two studies [ 103 , 107 ] had an unclear risk of detection bias and one of them [ 103 ] had a high risk of reporting bias. Another study [ 108 ] had a high risk of detection bias. Notwithstanding these higher levels of bias, these studies have been included in this review.

Summary of risk of bias in six studies.

Findings on outcome measurements

The outcome of the HAMD showed an overall decrease in the level of depression over time ( p <0.001) between the four treatment conditions (IPT, nefazodone, IPT and nefazodone, IPT and placebo), but this was not statistically significant. A significant difference was found between IPT and nefazodone and the use of nefazodone without IPT in favor of the first (for the intent to treat sample: adjusted OR (95% CI)=3.22 (1.02-10.12), p =0.045). Furthermore, a significant difference was found in the MADRS scores. Patients receiving IPT with nefazodone improved more than did patients receiving nefazodone without IPT. Furthermore, the nefazodone condition showed only a small improvement after the first six weeks [ 101 ].

Imipramine hydrochloride combined with clinical management (CM) was significantly superior to placebo with CM on general level of functioning. Patients receiving IPT or imipramine hydrochloride with CM appeared to have a better outcome on the HRSD than patients receiving placebo with CM ( p =0.018 and p =0.017). Furthermore, these patients showed a significantly higher percentage in the recovery analysis compared to placebo with CM patients, measured by a score of six or lower on the HRSD ( p =0.010 and p =0.013) [ 107 ].

In the Luty et al. study [ 102 ], depressive symptoms improved for about 55%. No statistically significant differences were found between IPT and CBT on the primary outcome measure MADRS (9.5% mean difference (95% CI), p =0.059), nor after controlling for baseline severity ( p =0.055).

HRSD scores were significantly higher in the IPT condition compared to the PHT-CM condition ( t (96)=−2.46, p <0.05, d =−0.50). No significant differences were found between IPT and CBT conditions ( t (96)=−1.19, p =0.46, d =−0.24), or between CBT and PHT-CM conditions ( t (96)=−1.35, p =0.37, d =−0.28) [ 103 ].

Depressive symptoms, measured by the HAMD and BDI, improved significantly ( p <0.001) in the first six weeks for patients receiving IPT or venlafaxine [ 104 ]. Although the venlafaxine group showed a slightly better outcome than the IPT group, no significant differences were found after six weeks.

O’Hara described recovery rates for women with PPD based on HRSD scores and BDI scores, favoring IPT over wait list condition (WLC). Based on HRSD scores (HRSD ≤6), IPT had a recovery rate of 31.7%, compared to 15% of WLC ( p =0.03). Based on BDI scores (BDI ≤9), IPT had a recovery rate of 38.3%, while women in the WLC group showed a recovery rate of 18.3% ( p =0.02) [ 108 ].

In both the IPT and CBASP group [ 105 ], HRSD scores decreased after 16 weeks, but only in the CBASP group statistical significance was reached ( t (13)=3.53, p =0.004). BDI scores were significantly lower after 16 weeks in both groups (IPT: t (14)=2.34, p =0.034; CBASP: t (13)=5.01, p <0.001). HRSD scores did not show a significant difference between the groups, whereas BDI scores showed a significantly higher reduction in depressive symptoms in the CBASP group after 16 weeks (mean BDI score of 10.79 vs. 21.27 in IPT; F (1,26)=4.34, p =0.047, treatment effect size: Cohen’s d =0.87).

Eight months after the start of the treatments (IPT, nortriptyline, or usual care), all HRSD scores improved significantly (χ 2 =816.14, df =6, p <0.001), and a significant difference was found between the groups (χ 2 =14.92, df =2, p =0.001). Post-hoc group t -test comparisons showed significant differences ( p <0.01) in HRSD scores between nortriptyline and usual care, at most measurement times favoring nortriptyline, and between IPT and usual care, favoring IPT after eight months. No significant difference was found between IPT and nortriptyline at any moment in time [ 106 ].

Meta-analysis and summary of findings

As can be seen in Table 2 , heterogeneity between the studies exists, which made it difficult to make meta-analytic comparisons. However, three studies were comparable in terms of measuring the effects of IPT and CBT [ 102 , 103 , 107 ]. The mean difference between the treatments was 1.01 (95% CI: -0.34, 2.37) favoring CBT over IPT, but did not reach a statistically significant level. See Figure 3 for more detailed information.

Comparison of HRSD scores between IPT and CBT. CBT Cognitive Behavior Therapy; IPT Interpersonal Psychotherapy; SD Standard Deviation.

Although no further meta-analyses were possible, and results appeared to be inconsistent, some conclusions can be drawn from these studies. IPT combined with nefazodone improved MADRS scores significantly better than did nefazodone alone [ 101 ]. Furthermore, higher HRSD scores were found in IPT patients in comparison with PHT-CM patients [ 103 ]. IPT patients showed a significantly greater decrease of HRSD and BDI scores than WLC patients [ 108 ]. As measured with the BDI, depressive symptoms were reduced more in CBASP patients in comparison with IPT patients [ 105 ]. Finally, IPT patients produced lower HRSD scores in comparison with patients receiving usual care [ 106 ].

Main results

The results of this systematic review show inconsistent findings in the eight heterogeneous studies included. The effectiveness and efficacy of the several treatments is comparable in most studies, and some conclusions may be drawn. Overall, the efficacy of IPT and CBT appears to be equal [ 102 ]. Contradictory results were found in IPT in comparison with pharmacotherapy. IPT combined with nefazodone appears to have a higher efficacy than sole nefazodone [ 101 ], while pharmacotherapy combined with clinical management appears to have a higher efficacy than IPT alone [ 103 ]. However, another study showed comparable results between IPT and imipramine hydrochloride with clinical management (CM), which both returned a better outcome on the HRSD compared to placebo with CM [ 107 ]. Furthermore, venlafaxine seems to reduce depressive symptoms more than IPT after six weeks, although this outcome was not significant [ 104 ]. The effects of using sole IPT and sole nortriptyline do not significantly differ from each other [ 106 ]. IPT and CBASP appear to be very comparable in efficacy, although scores of the BDI showed a slight preference for CBASP [ 105 ]. Finally, IPT appears to be more effective than wait list condition [ 108 ], and usual care after eight months, as does nortriptyline [ 106 ].

These outcomes suggest that several kinds of treatments are effective or efficacious for depressed patients, although one has to keep in mind the small number of included studies. Patients are recommended to choose a treatment which fits their personal preferences, since this may affect the outcome of the treatment. Policy makers are advised to base regulations on the effectiveness and efficacy of treatments in general, instead of a slightly different effect between one treatment and the other, since these studies do not take individual differences and preferences into account.

Limitations

This review has a number of limitations. First, this review included only adult outpatients with unipolar, non-psychotic major depression as a primary diagnosis. Although these inclusion criteria were a deliberate choice, this review has consequences for the generalizability. These results are not generalizable to children, adolescents, or the elderly, to patients with other kinds of depression, or to patients suffering from a combination of depression and medical conditions, or from depression and substance abuse. Furthermore, no distinction has been made in the severity of depression, which causes a higher heterogeneity in the complete sample, making results more uncertain.

Second, only eight studies with a limited number of participants were included in this review. Although most studies showed a low risk of bias, the small size of the sample may increase this risk. Furthermore, results are harder to generalize with a small number of participants, especially because many different kinds of treatments have been compared with each other (high heterogeneity), which limited the number of participants in the groups not receiving IPT. Moreover, the limited number of included studies in this review, makes one question the applicability of the Cochrane guidelines for conducting a systematic review [ 52 ], for clinical treatments in mental health care.

Third, all included patients were outpatients and therefore had to be willing and motivated to participate in the selected studies. This may cause some bias, since not all types of patients could be included in the studies. For example, treatment-resistant depressed patients may have been less motivated than patients who were not treatment-resistant, and it may not be possible to generalize results for these patients.

Fourth, pharmacotherapy consisted of different types of antidepressant medication. Although these medications may seem to be equally effective, some differences may exist, which may interfere with the results of this review. Furthermore, one study [ 101 ] used nefazodone as pharmacotherapy, although this medication has been withdrawn in, amongst other countries, the USA and the Netherlands, because of hepatotoxicity associated with this drug [ 114 ].

Fifth, some of the findings were based on the scores of the HRSD [ 101 , 103 – 108 ]. However, this scale has recently been criticized for having multiple problems, including among others the existence of different versions and not being as sensitive as other scales [ 115 , 116 ]. Despite these flaws, the HRSD has been used in many studies and the outcomes of this scale can therefore not be excluded from this review. Furthermore, findings were also based on the MADRS [ 101 , 102 ], which is more sensitive to treatment effect than the HAMD [ 117 ], and on the BDI [ 104 , 105 ] which correlates weakly with the HDRS [ 118 ] and has several advantages and disadvantages [ 119 ], but is widely used.

Sixth, one study [ 104 ] measured the efficacy only after six weeks, without follow-up measurement. This is a very short period for measuring the efficacy of IPT. Therefore, the results of this study may be questionable. Furthermore, these authors did not include an intention-to-treat analysis, which increases the risk of bias.

Finally, although a profound search has been performed, there is no complete certainty that all studies eligible for this review have been found. Furthermore, the search was directed only at published studies, automatically excluding unpublished data, causing possible publication bias.

It can be concluded that the differences between the effects and efficacy of several types of treatment are very small and they are often not significant. This in turn is consistent with a study concluding that the effects of psychotherapy for adult depression in meta-analyses are overestimated [ 27 ]. Nevertheless, usual care, as described in the study of Schulberg et al. [ 106 ], appears to be ineffective and is not recommended as a treatment for MDD. Therefore, psychotherapeutic treatments such as IPT and CBT, and/or pharmacotherapy are recommended as first-line treatments for depressed adult outpatients. This conclusion is consistent with a previous study [ 21 ], and review [ 26 ], and previous meta-analyses [ 28 , 29 , 32 , 33 , 55 ],although, as has been stated in the introduction, these studies had several limitations as well. Furthermore, it is recommended that the type of treatment is adjusted to the individual preferences of the patient.

Future research should focus on a larger sample including patients with MDD, while correcting for severity of depression. Since many studies focused on IPT combined with medication, it is recommended that these studies be included in future research as well. Furthermore, it is recommended that future studies included in a review, have longer follow-up periods. All studies should aim for the highest quality standards currently set.

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This study was not funded by any grants. We thank Tim Ellermann and Henrietta Hazen for help during the development of an adequate search strategy. MH also thanks SE and TR for their support.

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MH designed the study with the support of SE and TR. MH undertook the literature search with help from TE, identified potential and final selected articles, interpreted results, drafted and revised all versions of the manuscript, supported by SE and TR. In case of doubt during the screening and analyzing phase, TR was consulted. SE and TR supervised the development of the manuscript. All authors read and approved the final version.

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van Hees, M.L.J.M., Rotter, T., Ellermann, T. et al. The effectiveness of individual interpersonal psychotherapy as a treatment for major depressive disorder in adult outpatients: a systematic review. BMC Psychiatry 13 , 22 (2013). https://doi.org/10.1186/1471-244X-13-22

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The efficacy of psychotherapies and pharmacotherapies for mental disorders in adults: an umbrella review and meta‐analytic evaluation of recent meta‐analyses

Falk leichsenring, christiane steinert, sven rabung, john pa ioannidis.

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Mental disorders represent a worldwide public health concern. Psychotherapies and pharmacotherapies are recommended as first line treatments. However, evidence has emerged that their efficacy may be overestimated, due to a variety of shortcomings in clinical trials (e.g., publication bias, weak control conditions such as waiting list). We performed an umbrella review of recent meta‐analyses of randomized controlled trials (RCTs) of psychotherapies and pharmacotherapies for the main mental disorders in adults. We selected meta‐analyses that formally assessed risk of bias or quality of studies, excluded weak comparators, and used effect sizes for target symptoms as primary outcome. We searched PubMed and PsycINFO and individual records of the Cochrane Library for meta‐analyses published between January 2014 and March 2021 comparing psychotherapies or pharmacotherapies with placebo or treatment‐as‐usual (TAU), or psychotherapies vs. pharmacotherapies head‐to‐head, or the combination of psychotherapy with pharmacotherapy to either monotherapy. One hundred and two meta‐analyses, encompassing 3,782 RCTs and 650,514 patients, were included, covering depressive disorders, anxiety disorders, post‐traumatic stress disorder, obsessive‐compulsive disorder, somatoform disorders, eating disorders, attention‐deficit/hyperactivity disorder, substance use disorders, insomnia, schizophrenia spectrum disorders, and bipolar disorder. Across disorders and treatments, the majority of effect sizes for target symptoms were small. A random effect meta‐analytic evaluation of the effect sizes reported by the largest meta‐analyses per disorder yielded a standardized mean difference (SMD) of 0.34 (95% CI: 0.26‐0.42) for psychotherapies and 0.36 (95% CI: 0.32‐0.41) for pharmacotherapies compared with placebo or TAU. The SMD for head‐to‐head comparisons of psychotherapies vs. pharmacotherapies was 0.11 (95% CI: –0.05 to 0.26). The SMD for the combined treatment compared with either monotherapy was 0.31 (95% CI: 0.19‐0.44). Risk of bias was often high. After more than half a century of research, thousands of RCTs and millions of invested funds, the effect sizes of psychotherapies and pharmacotherapies for mental disorders are limited, suggesting a ceiling effect for treatment research as presently conducted. A paradigm shift in research seems to be required to achieve further progress.

Keywords: Psychotherapies, pharmacotherapies, mental disorders, randomized controlled trials, meta‐analyses, effect sizes, meta‐analytic evaluation

Mental disorders represent a worldwide public health concern 1 , 2 . Psychotherapies and pharmacotherapies are recommended as first line treatments 3 , 4 . However, evidence has recently emerged suggesting that the efficacy of both types of treatment may have been overestimated, due to several shortcomings of clinical trials, such as publication bias, researcher allegiance, or use of weak comparison groups (in particular, waiting list) 5 , 6 , 7 , 8 , 9 , 10 , 11 , 12 , 13 , 14 , 15 , 16 . A realistic estimate of the efficacy of psychotherapies and pharmacotherapies is important to obtain.

Meta‐analyses and systematic reviews of randomized controlled trials (RCTs) are thought to provide the highest level of evidence 17 . However, not only individual RCTs but also meta‐analyses may be affected by the above‐mentioned biases 6 , 18 , 19 . To avoid overestimating treatment efficacy, meta‐analyses need to take risk of bias systematically into account 6 , 18 , 19 , 20 . Furthermore, estimates of efficacy depend upon the comparator against which treatments are tested. Waiting list conditions, for example, represent a relatively weak comparator, leading to larger effect sizes 6 , 8 , 21 , 22 . Comparisons with placebo or treatment‐as‐usual (TAU) provide better estimates of the true efficacy of a treatment 6 , 22 .

The most recent comprehensive review of meta‐analyses of both psychotherapies and pharmacotherapies in mental disorders, including 61 meta‐analyses, was published in 2014, reporting a medium effect size (standardized mean difference, SMD = 0.50) 8 . Some of the included meta‐analyses, however, used waiting list comparisons in the assessment of overall efficacy. In addition, the authors seem to have just averaged the extracted effect sizes, without performing a meta‐analytic evaluation including weighting effect sizes. Furthermore, a large number of studies and meta‐analyses have been published since 2014.

For all these reasons, we carried out an up‐to‐date umbrella review of recent meta‐analyses of psychotherapies and pharmacotherapies for the main mental disorders in adults which used placebo or TAU as comparison groups and formally assessed risk of bias or quality of studies. As the primary outcome, we used the effect size for target symptoms of the relevant disorder.

The study protocol of this umbrella review was registered in advance at PROSPERO (International Prospective Register of Systematic Reviews), registration number: CRD42020155452.

Inclusion criteria

Meta‐analyses of RCTs comparing psychotherapies or pharmacotherapies to placebo or TAU in adults with mental disorders published since 2014 were eligible. We also considered meta‐analyses comparing psychotherapies vs. pharmacotherapies head‐to‐head, or their combination to either monotherapy. Only meta‐analyses which formally assessed risk of bias or quality of studies were included. If multiple meta‐analyses fulfilling the inclusion criteria were available for one condition, all of them were included. Reporting of SMD or other measures of between‐group effect size was required.

All types of pharmacotherapy or psychotherapy were eligible for inclusion. Meta‐analyses examining specific subgroups (e.g., treatment resistant depression, primary care patients, the elderly), psychiatric or somatic comorbidities (e.g., depression in cancer patients), specific settings (e.g., group therapy only, or inpatient therapy) or augmentation strategies (e.g., psychostimulants added to antipsychotic drugs in schizophrenia), or focusing on secondary outcomes (e.g., quality of life in depression) were not included. These inclusion criteria are consistent with the above‐mentioned 2014 review 8 , except for excluding waiting list comparisons and requiring meta‐analyses to have assessed risk of bias or quality of studies. Both standard and network meta‐analyses were eligible.

Combining data of patients receiving TAU or placebo with those of patients on waiting list has been shown to inflate effect sizes 8 , 22 , 23 . On the other hand, mixing data of patients on TAU with those receiving specific therapies (e.g., cognitive‐behaviour therapy, CBT) can be expected to underestimate the effect size of the treatment in question. Therefore, meta‐analyses mixing data of TAU or placebo with waiting list or active treatments were excluded.

Search for studies

We searched PubMed and PsycINFO and individual records of the Cochrane Library for meta‐analyses of RCTs of psychotherapies and/or pharmacotherapies for mental disorders in adults published between January 2014 and March 2021.

Four reviewers independently searched for studies. Decision on inclusion was made by consensus including another rater. Search terms were meta‐analy* or metaanaly* combined with the thesaurus of the individual databases concerning each disorder. To provide comparable results, we used the syntax applied in the previous most comprehensive review 8 .

Data extraction

We focused on effect sizes and 95% confidence intervals (CIs) for the target symptoms of the relevant disorder (primary outcome). We extracted between‐group SMDs and related measures (Cohen's d, Hedges' g) as reported in the meta‐analyses. Odds ratios (ORs) and hazard ratios (HRs) were converted to SMDs 24 , 25 . Data on relative risk (RR) were extracted as reported. We used Cohen's convention of d=0.2, d=0.5 and d=0.8 as indicating small, medium and large effect sizes 26 , corresponding to success rate differences of 11%, 28% and 43%; numbers needed to treat of 9, 4 and 2; ORs of 1.43, 2.48 and 4.25; RRs of 1.22, 1.86 and 3.00; and HRs of 1.3, 1.9 and 2.8 24 , 25 , 27 , 28 , 29 . Intention‐to‐treat data were preferred whenever available.

If meta‐analyses took risk of bias into account by, for example, additionally reporting data separately for low risk of bias studies or correcting for publication bias, we listed all reported effect sizes but preferably focused on the corrected or high quality data for interpreting results.

Rates of remission and response were included as secondary outcomes when available. Dichotomous variables have some limitations 30 , but complementarily to SMDs they can provide useful information about efficacy.

One author extracted data (type of treatment and disorder, number of studies, number of participants, type of comparator, risk of bias, adverse events/side effects, and effect sizes). Data were cross‐checked independently by two raters each.

Quality assessment

The quality of the included meta‐analyses was independently assessed by two raters. For the purpose of this review, we used the items 1 to 9 of the Checklist for Systematic Reviews and Research Syntheses 31 , 32 , complemented by item 12 of AMSTAR 2 20 (“Was the impact of risk of bias in individual studies on results of the meta‐analysis taken into account?”) and an additional item addressing whether the meta‐analysis was registered. In case of disagreement between raters, consensus ratings were used.

Data synthesis

The results of the largest meta‐analyses for each condition, i.e. those including most RCTs, are presented and evaluated separately. Additionally, these independent meta‐analyses were included in second‐order meta‐analyses combining their summary effect sizes across all the different mental disorders 33 . This allowed to obtain a weighted effect of psychotherapy or pharmacotherapy across all mental disorders, and weighted effects for the benefits of combined therapy, and for the comparative efficacy of psychotherapy vs. pharmacotherapy. The analysis was performed by Comprehensive Meta‐Analysis (CMA, Version 3) using a random effects model based on SMDs and their CIs via the CMA analysis option “generic estimates”.

Heterogeneity was assessed using the I 2 statistic. If meta‐analyses did not report an overall effect size, but effect sizes for specific treatments and comparisons, the effect sizes of the relevant comparisons were aggregated by CMA and the resulting overall SMDs were entered into the second‐order meta‐analyses across disorders. Only effect estimates based on at least two primary RCTs were used.

Included meta‐analyses

The search retrieved 23,601 items, reduced to 19,500 after removing duplicates, which were screened by titles and abstracts. Full‐text evaluation was carried out for 319 papers. One hundred and two meta‐analyses fulfilled the inclusion criteria (see Figure 1 and supplementary information). These encompassed 69 meta‐analytic comparisons of pharmacotherapies with placebo or TAU, 26 comparisons of psychotherapies with placebo or TAU, 11 comparisons of psychotherapies vs. pharmacotherapies head‐to‐head, and 13 comparisons of combined psychotherapy and pharmacotherapy to either monotherapy 6 , 12 , 13 , 34 , 35 , 36 , 37 , 38 , 39 , 40 , 41 , 42 , 43 , 44 , 45 , 46 , 47 , 48 , 49 , 50 , 51 , 52 , 53 , 54 , 55 , 56 , 57 , 58 , 59 , 60 , 61 , 62 , 63 , 64 , 65 , 66 , 67 , 68 , 69 , 70 , 71 , 72 , 73 , 74 , 75 , 76 , 77 , 78 , 79 , 80 , 81 , 82 , 83 , 84 , 85 , 86 , 87 , 88 , 89 , 90 , 91 , 92 , 93 , 94 , 95 , 96 , 97 , 98 , 99 , 100 , 101 , 102 , 103 , 104 , 105 , 106 , 107 , 108 , 109 , 110 , 111 , 112 , 113 , 114 , 115 , 116 , 117 , 118 , 119 , 120 , 121 , 122 , 123 , 124 , 125 , 126 , 127 , 128 , 129 , 130 , 131 , 132 , 133 , 134 . The 102 meta‐analyses encompassed 3,782 RCTs (range: 2 to 522) and 650,514 patients (range: 65 to 116,477) (see supplementary information).

PRISMA flow chart. RoB – risk of bias, TAU – treatment as usual

Across all meta‐analyses, the mean number of positively rated items in the quality assessment was 8.71±1.43 (range: 4 to 11). The items 10 (item 12 of AMSTAR 2, addressing whether the meta‐analyses took the impact of bias on results into account) and 11 (study registration) were the least frequently fulfilled (48% and 47%, respectively). The quality of meta‐analyses was not significantly different between psychotherapies and pharmacotherapies (mean of positively rated items: 8.95±1.12 for psychotherapies, 8.68±1.54 for pharmacotherapies, t=0.74, p=0.46).

Psychotherapies and pharmacotherapies vs. TAU or placebo

In the largest meta‐analyses, the effect sizes of both psychotherapies and pharmacotherapies in comparison to TAU or placebo were small (SMD<0.50) for most disorders and treatments (see Figure 2 and supplementary information). Medium effect sizes were found only for pharmacotherapies of obsessive‐compulsive disorder (OCD) (SMD=0.56) 72 , bulimia nervosa (SMD=0.61) 80 , and somatoform disorders (SMD=0.50) 91 , and for psychotherapies of post‐traumatic stress disorder (PTSD) (SMD=0.54) 54 and borderline personality disorder (SMD=0.57) 93 . Large effect sizes were only reported for psychotherapy of OCD (SMD=1.03) 74 , with, however, a substantial proportion of patients taking concomitant pharmacotherapy 72 , 74 . Overall, risk of bias was often high (see Figure 2 and supplementary information).

Effect sizes in the largest meta‐analyses of pharmacotherapies (squares) and psychotherapies (circles) in comparison to placebo or treatment‐as‐usual (TAU). PHA – pharmacotherapy; PSY – psychotherapy, SMD – standardized mean difference, * – adjusted for risk of bias, ° – adjusted for small‐study effects, MDD – major depressive disorder, GAD – generalized anxiety disorder, SAD – social anxiety disorder, OCD – obsessive‐compulsive disorder, PTSD – post‐traumatic stress disorder, PD – personality disorder, ADHD – attention‐deficit/hyperactivity disorder, H – high, M – medium, L – low, U – uncertain, NR – not reported. Where SMD is not provided, this means that no valid meta‐analysis reporting this value was available.

For psychotherapies and pharmacotherapies, second‐order random effects meta‐analyses in comparison to placebo or TAU yielded statistically significant but small SMDs of 0.34 (95% CI: 0.26‐0.42, I 2 =66.33%) and 0.36 (95% CI: 0.32‐0.41, I 2 =70.61%), respectively, across disorders (see Figure 2 ). For the aggregated data of psychotherapies and pharmacotherapies, the SMD was 0.35 (95 CI: 0.31‐0.39, I 2 =68.23%).

Depressive disorders

For psychotherapies of depressive disorders, the largest meta‐analysis reported a small SMD of 0.31, adjusted for biases, in comparison to TAU 51 (see Figure 2 ). Taking all included meta‐analyses into account, psychotherapy achieved effect sizes (SMDs) between 0.11 and 0.61 in comparison to placebo or TAU 6 , 12 , 37 , 50 , 51 , 52 , except for one outlying meta‐analysis reporting a large SMD post‐therapy (1.11), reduced to 0.27 at 3 to 12 month follow‐up and associated with a high risk of bias 52 . The majority of effect sizes were small (<0.50).

Only between 1% and 17% of studies of psychotherapy for depression were found to show a low risk of bias. When meta‐analyses took risk of bias into account, they consistently found a decrease in effect sizes (see supplementary information).

Across all forms of psychotherapy, remission from major depressive disorder (Hamilton Depression Rating Scale, HAM‐D <7) was achieved in 43% of patients, with no significant differences between the various psychotherapies 5 . Response (50% reduction of HAM‐D score) was achieved in 54% of patients 5 . TAU was superior to no treatment with regard to remission (33% vs. 23%), but inferior to psychotherapy (33% vs. 43%) 135 .

The largest meta‐analysis of pharmacotherapy for depressive disorders reported a SMD of 0.30 36 (see Figure 2 ). All effect sizes (SMD) achieved by pharmacotherapy in comparison to placebo were below 0.50, ranging from 0.19 to 0.41. The exception was ketamine, which achieved large short‐term effects (0.83, 0.88) 24 hours and 3‐4 days after treatment, dropping to 0.31 after 7 days 13 , 34 , 35 , 36 , 37 , 38 , 39 , 41 , 42 , 43 , 44 , 45 , 46 , 47 , 48 , 49 . Most effect sizes in terms of RRs were small as well (≤1.22).

The mean response rate for selective serotonin reuptake inhibitors (SSRIs) was 51% vs. 39% for placebo 35 , corresponding to a small effect size 27 .

Many trials of pharmacotherapy in depression showed a high risk of bias 13 , 35 , 36 , 42 (see supplementary information).

Anxiety disorders

In the largest meta‐analyses of anxiety disorders, psychotherapies achieved SMDs between 0.28 and 0.44 6,55,66 (see Figure 2 ). Overall, psychotherapies of anxiety disorders achieved SMDs compared to TAU or placebo between 0.01 and 0.72 6 , 54 , 55 , 59 , 65 , 66 , 71 , except for two outlying effect sizes in generalized anxiety disorder (1.44, 1.32), each based on three studies only 6 , 55 . Two effect sizes of psychotherapy (CBT) in social anxiety disorder were medium (0.72, 0.56) 66 , but most effect sizes were small (see supplementary information).

Overall, only 17% of psychotherapy studies in anxiety disorders were found to show a low risk of bias 6 .

In the largest meta‐analyses for anxiety disorders, pharmacotherapies achieved SMDs in comparison to placebo between 0.33 and 0.45 53,56,64 (see Figure 2 ). Overall, effect sizes for pharmacotherapy were between 0.01 and 0.56, with the majority of effect sizes being small 53 , 55 , 56 , 57 , 58 , 59 , 60 , 61 , 62 , 63 , 64 , 66 , 67 , 69 , 70 (see supplementary information). RR ranged between 1.20 and 4.03, with most values being small, one medium (monoamine oxidase inhibitors), and one large (benzodiazepines, RR=4.03) 69 .

For social anxiety disorder and generalized anxiety disorder, pharmacotherapy yielded response rates of 52% and 56%, respectively, versus 32% and 41% with placebo 59 , 69 .

Obsessive‐compulsive disorder

For psychotherapy (CBT) of OCD, the largest meta‐analysis reported a large SMD (1.03) 74 (see Figure 2 ). Considering all meta‐analyses, large SMDs in comparison to placebo were reported (0.91‐1.46) 72 , 74 . At follow‐up of on average of 15.1 months after the end of treatment, SMDs decreased from 0.57 to 0.06 for all comparators 74 . Follow‐up results were not available for a comparison against placebo. Most psychotherapy trials included patients taking stable doses of antidepressants 72 , 74 , possibly overestimating effect sizes in favour of psychotherapy 72 .

For pharmacotherapy of OCD, the largest meta‐analysis reported a medium effect size (SMD=0.56) 72 (see Figure 2 ). Considering all meta‐analyses, small to medium SMDs were reported (0.45‐0.66).

For most studies of psychotherapy and pharmacotherapy, the risk of bias was high (see Figure 2 and supplementary information).

Post‐traumatic stress disorder

For psychotherapy (CBT) of PTSD, the largest meta‐analysis reported a medium effect size compared to TAU (SMD=0.54) 54 (see Figure 2 ), which was stable at follow‐ups of up to 12 months after end of therapy 54 . For PTSD related to childhood maltreatment, a SMD of 0.50 in comparison to TAU/placebo was found, which was reduced to 0.21 after adjusting for small sample size 79 .

For pharmacotherapy of PTSD, the largest meta‐analysis reported a small SMD in comparison to placebo (0.21) 76 (see Figure 2 ). Considering all meta‐analyses, effect sizes achieved by pharmacotherapy in comparison to placebo were heterogeneous (SMDs: –0.10 to 0.97) 75 , 76 , 77 , 78 . Risk of bias was often high 77 , 78 . A large SMD was obtained with phenelzine (0.97), a medium one with mirtazapine (0.79), desipramine (0.52) and olanzapine (0.51), all based on only one RCT except for olanzapine 75 . For SSRIs and serotonin and norepinephrine reuptake inhibitors (SNRIs), a medium SMD was reported (0.50) 77 . For all other drugs, SMDs were <0.50 (from –0.10 to 0.47).

Personality disorders

For psychotherapy of personality disorders, only a meta‐analysis of borderline personality disorder was available, which reported a medium SMD in comparison to TAU (0.57), with a high risk of bias in most studies (see Figure 2 ) 93 .

An update for developing a Cochrane report of pharmacotherapy in borderline personality disorder did not provide meta‐analytic results since the authors did not find robust evidence 136 .

Somatoform disorders

For psychotherapy of somatoform disorders, the largest meta‐analysis reported a small SMD (0.19, see Figure 2 ) in comparison to enhanced care, with high risk of bias due to lack of blinding 92 . For pharmacotherapy of somatoform disorders, the largest meta‐analysis reported a medium SMD (0.50, see Figure 2 ) in comparison with placebo 91 .

Considering all meta‐analyses, heterogeneous SMDs (from 0.13 to 0.91) were reported for pharmacotherapy, based on two or three RCTs, with a high risk of bias for most RCTs 91 .

Eating disorders

For psychotherapy of bulimia nervosa, no recent meta‐analysis fulfilled the inclusion criteria. For pharmacotherapy, the largest meta‐analysis reported a medium SMD in comparison with placebo (0.61, see Figure 2 ) 80 . Considering all meta‐analyses, considerable heterogeneity among classes of drugs were found (SMDs: 0.10‐1.00) 80 .

For psychotherapy of binge eating disorder, no meta‐analysis fulfilled the inclusion criteria. For pharmacotherapy, the largest meta‐analysis reported a small to medium SMD in comparison with placebo (0.45, see Figure 2 ) 82 . Considering all meta‐analyses, a small to medium effect size compared to placebo was found for pharmacotherapy (SMD=0.45, RR: 1.67, 2.61) 81 , 82 . One of these meta‐analyses reported a high 82 , the other a medium to low risk of bias 81 .

For psychotherapy of anorexia nervosa, the largest meta‐analysis reported a small SMD in comparison with TAU (0.14, see Figure 2 ) 85 . Overall, the effect sizes in comparison to TAU or placebo were small (SMD=0.10‐0.31, RR: 0.97, 1.28) 84 , 85 , 86 . For pharmacotherapy, the largest meta‐analysis reported a small effect size (SMD=0.25) 83 .

Substance use disorders

For both psychotherapy and pharmacotherapy of substance use disorders, the largest meta‐analysis reported small SMDs in comparison with TAU or placebo (0.23 and 0.26, respectively, see Figure 2 ) 95 , 96 . For psychotherapy, the effect size decreased at follow‐ups ≥8 months after end of treatment (SMD=0.05) 96 . Considering all meta‐analyses, small effect sizes were found for pharmacotherapy (SMDs: 0.07 to 0.35, RR: 0.32‐1.39) 94 , 95 .

For psychotherapy, no recent meta‐analysis fulfilled the inclusion criteria. The quality of studies was found to be low 137 . For pharmacotherapy, the largest meta‐analysis reported a small SMD in comparison with placebo (0.27, see Figure 2 ) 90 . Overall, for pharmacotherapy of insomnia, small to medium SMDs were reported (0.07 to 0.58) 88 , 89 , 90 . One meta‐analysis provided effect sizes only for one of eight outcome measures, with large SMDs (0.88‐1.38), suggesting selective reporting 87 .

Attention‐deficit/hyperactivity disorder (ADHD)

For psychotherapy of ADHD in adults, no meta‐analysis could be included 134 . For pharmacotherapy, the largest meta‐analysis reported a small to medium SMD in comparison with placebo (0.45, see Figure 2 ) 99 . Considering all meta‐analyses, the effects of pharmacotherapy were heterogeneous (from 0.16 to 0.97) 98 , 99 , 100 , 101 , 102 . Large SMDs were found for amphetamines 98 , 100 , 102 , small to medium SMDs for methylphenidate 100 , 101 . Risk of bias was often high or unclear, and level of evidence was low to very low 98 , 100 .

Schizophrenia spectrum disorders

Results of psychotherapy in schizophrenia spectrum disorders were evaluated in the context of pharmacotherapy (i.e., patients usually received concomitant medication). The largest meta‐analysis reported a small SMD in comparison with TAU (0.33, see Figure 2 ) 111 . Considering all meta‐analyses, small effect sizes compared to nonspecific controls were found for overall symptoms, positive and negative symptoms (SMDs: 0.32, 0.24 and 0.08, respectively) 138 . In comparison to TAU, psychotherapy achieved small to medium SMDs for negative symptoms (0.15‐0.58) 110 , 112 .

For psychotherapy, a response rate of 13% for overall symptoms and 25% for positive symptoms was found 139 (a reduction of symptoms of at least 50% was required). The response rate decreased considerably if researcher allegiance (authors evaluated the therapy that they developed) was taken into account (from 13% to 4.9%) 139 .

For acute pharmacological treatment of schizophrenia, the largest meta‐analysis reported an overall SMD of 0.45 for target symptoms, reduced to 0.38 after adjusting for publication bias (Figure 2 ) 103 . These results are consistent with meta‐analyses on specific drugs, such as quetiapine (SMD=0.33), cariprazine (SMD: 0.32‐0.37), lurasidone (SMD: 0.34‐0.47), and aripiprazole and brexpiprazole (RR=1.1) 104 , 107 , 108 , 109 . A large effect size was reported for clozapine (SMD=0.89) 103 . Large and medium SMDs were achieved by long‐acting injectable antipsychotics in the maintenance treatment of non‐affective psychoses (RR: 1.75‐3.70) 107 .

For the acute treatment of schizophrenia with pharmacotherapy, differences in response rates in comparison with placebo were small (23% vs. 14%) 14 .

Bipolar disorder

For psychotherapy of bipolar disorder, the largest meta‐analysis reported a small effect size in comparison to TAU (SMD=0.18, see Figure 2 ) 121 , with small effect sizes for both depression and mania symptoms (SMDs: 0.23 and 0.05, respectively) and for relapse prevention post‐therapy (RR: 1.52) 121 . At follow‐up 26 to 78 weeks post‐therapy, SMDs were 0.21 and 0.38; RR for relapse was 1.35 121 . Psychotherapy was given in the context of concomitant pharmacotherapy.

For the acute treatment of mania, the results for pharmacotherapy are heterogeneous. One meta‐analysis reported medium SMDs for cariprazine (0.51‐0.52) 119 , and another reported a small effect size for aripiprazole (SMD=0.16) 115 . These two meta‐analyses included only three RCTs. A third meta‐analysis reported a very large SMD (1.51) for tamoxifen 120 , based on two RCTs with small samples (16 and 66 cases, respectively), making the results questionable. This study represents a clear outlier.

For the acute treatment of bipolar depression, the largest meta‐analysis of pharmacotherapy reported heterogeneous results, with effect sizes (SMDs) between 1.41 and –1.84 113 . Large effect sizes were achieved by fluoxetine (1.41), divalproex (1.25), lurasidone (1.15), moclobemide (1.09), cariprazine (0.85) and imipramine (0.86) 113 , all of them based, however, on only 0‐3 direct comparisons. Some drugs achieved medium effect sizes (olanzapine, phenelzine, tranylcypromine). The effect sizes of all other drugs were small 113 . Quetiapine achieved an almost medium effect size (SMD=0.48) based on 11 direct comparisons 113 .

For the prevention of manic/hypomanic/mixed episodes, effect sizes of lithium were almost medium (RR=1.85) 114 . Medium effect sizes were reported for olanzapine (RR=2.88) and risperidone (RR=2.88); large effect sizes for aripiprazole (once monthly) and asenapine (RR: 3.31, 4.81) 114 . The results for asenapine are based on only one RCT 116 . For all other drugs, effect sizes were small 114 .

For the prevention of any mood episode relapse, a large effect size was found for asenapine (RR=3.82), with the caveat mentioned above 114 . Medium effect sizes were reported for quetiapine and olanzapine 114 , 116 , 118 . Small effect sizes for the prevention of any mood episode were achieved by lithium (RR: 1.60, 1.61) and several other drugs 114 , 118 . Earlier meta‐analyses reported heterogeneous results for the prevention of any relapse by lithium (SMD: 1.12, 0.47) 140 .

For the prevention of depressive episodes, quetiapine and asenapine achieved medium effect sizes (RR: 2.08, 2.60). For all other drugs, including lithium (RR=1.26), effect sizes were small 114 . Small effect sizes were found for antidepressants (RR=1.56) 117 .

Psychotherapies vs. pharmacotherapies

Head‐to‐head comparisons of psychotherapies vs. pharmacotherapies yielded small effect sizes for all disorders examined, i.e., depressive disorders, anxiety disorders, PTSD and OCD (SMDs: 0.00‐0.24, see Figure 3 ) 37 , 66 , 74 , 126 . A second‐order random effects meta‐analysis across the effect sizes of the largest meta‐analyses (Figure 3 ) yielded a non‐significant SMD of 0.11 (95% CI: –0.05 to 0.26, I 2 =61.99).

Effect sizes in the largest meta‐analyses for head‐to‐head comparisons of psychotherapies (PSY) vs. pharmacotherapies (PHA). SMD – standardized mean difference, ° – adjusted for small‐study effects, MDD – major depressive disorder, SAD – social anxiety disorder, OCD – obsessive‐compulsive disorder, PTSD – post‐traumatic stress disorder, H – high, M – medium, L – low, U – uncertain, NR – not reported

Considering all included meta‐analyses, no substantial differences in short‐term efficacy between psychotherapies and pharmacotherapies in depressive disorders, anxiety disorders and PTSD were found 12 , 37 , 66 , 122 , 123 , 124 , 125 , 126 , with only a few exceptions. In OCD, psychotherapy achieved medium to large SMDs (0.61‐0.95) in comparison to SSRIs 72 , but most psychotherapy trials included patients taking stable doses of antidepressants, affecting results in favour of psychotherapy. Most studies of psychotherapy and pharmacotherapy in OCD had a high risk of bias 72 . With regard to long‐term efficacy, psychotherapy achieved a large SMD compared to pharmacotherapy in PTSD (0.83) 126 . For other disorders, no head‐to‐head comparisons fulfilled the inclusion criteria.

Combining psychotherapy and pharmacotherapy

In the largest meta‐analyses (Figure 4 ), effect sizes (SMDs) in favour of the combined treatment were small for most disorders, that is depressive disorders (0.37, 0.15) 37 , social anxiety disorder (combined vs. pharmacotherapy: 0.40) 66 , OCD (combined vs. psychotherapy: 0.25) and PTSD (0.09, 0.12) 126 . Effect sizes (SMDs) were medium in favour of the combined treatment vs. psychotherapy in social anxiety disorder (0.52) 66 and for the combined treatment vs. pharmacotherapy (SSRIs) in OCD (0.73), based on a network meta‐analysis including only one direct comparison with very small samples 72 . A large effect size was found only for the combined treatment vs. pharmacotherapy in ADHD (0.80) 134 , based on only two RCTs showing a high risk of bias in at least one domain.

Effect sizes in the largest meta‐analyses for combined therapy vs. pharmacological (squares) or psychological (circles) monotherapy. SMD – standardized mean difference, ° – adjusted for small‐study effects, COM – combined therapy, PHA – pharmacotherapy, PSY – psychotherapy, MONO – monotherapy, MDD – major depressive disorder, SAD – social anxiety disorder, OCD – obsessive‐compulsive disorder, PTSD – post‐traumatic stress disorder, ADHD – attention‐deficit/hyperactivity disorder, H – high, M – medium, L – low, U – uncertain, NR – not reported

A second‐order random effects meta‐analysis across the effect sizes of the largest meta‐analyses yielded a statistically significant but small SMD of 0.31 (95% CI: 0.19‐0.44, I 2 =53.02) in favour of the combined treatment (Figure 4 ).

Considering all included meta‐analyses, most effect sizes (SMDs) achieved by the combined treatment compared to either monotherapy in depressive disorders, anxiety disorders, PTSD, OCD and ADHD were small (0.09‐0.48) when risk of bias was taken into account 12 , 37 , 66 , 72 , 74 , 126 , 128 , 129 , 131 , 132 , 134 . Exceptions were the superiority of the combined treatment in long‐term outcome of PTSD over pharmacotherapy (SMD=0.96, based on only two direct comparisons) 126 , and the superiority of the combined treatment over psychodynamic therapy in social anxiety disorder (SMD=0.68), based on a network meta‐analysis including zero direct comparisons for the condition 66 , making a study of inconsistencies impossible 141 .

In several of these meta‐analyses, risk of bias was high in several domains, or results were based on only a few or small studies 66 , 72 , 126 , 134 .

In this field‐wide assessment of psychotherapies and pharmacotherapies for mental disorders in adults, we included evidence from 102 meta‐analyses with 3,782 RCTs and 650,514 patients. We found small benefits overall for both types of interventions, with an average SMD of 0.35 and moderate heterogeneity across conditions 142 . This finding challenges the result of the previous most comprehensive review, which reported an overall medium effect size (SMD=0.50) across psychotherapies and pharmacotherapies, based on 61 meta‐analyses with 852 RCTs and 137,126 patients 8 . This latter estimate seems to be due to including waiting list comparators and averaging effect sizes without performing a random effects meta‐analytic evaluation.

According to the results of this umbrella review and second‐order meta‐analyses, there is an additional gain of psychotherapy and pharmacotherapy in the treatment of mental disorders in adults, but this is small in terms of effect sizes 26 . Conditions for which very extensive evidence was available (e.g., depression) almost always had such modest effect sizes when only studies with low risk of bias were considered, or efforts were made to correct for bias. Medium or large effect sizes were found only for few conditions, and most of the effects sizes ≥0.50 were associated with a high risk of bias and/or limited evidence. Nevertheless, the argument still holds that, although there are some medications for general medical conditions with clearly higher effect sizes, psychotropic agents or psychotherapies are not generally less efficacious than those medications 140 .

Some limitations and features of this umbrella review should be discussed as they affect the interpretation of overall evidence. First, several meta‐analytic comparisons included only a few studies, affecting statistical power and external validity of results.

Second, the results of network meta‐analyses need some extra caution 141 , 143 . It has been argued that these meta‐analyses can only provide observational evidence, since the comparisons between treatments are both direct and indirect, and the latter are not randomized 144 . As a related issue, transitivity (similar distribution of effect modifiers) can be controlled statistically only for known modifiers, in contrast to controlling all modifiers by randomization. Some of the network meta‐analyses included in this review encompassed only a few or even no direct comparisons of specific treatments 66 , 72 , 75 . Statistical power may be low if only a few studies with small samples and large heterogeneity are included 141 , 145 . Thus, some inconsistencies between direct and indirect comparisons may not have been detected 145 , possibly affecting effect size estimates.

Third, we followed Cohen's convention of small, medium and large effect sizes 26 . However, the clinical relevance of these estimates is not clear. The clinical benefit of an intervention needs to be determined by comparison with a benchmark such as the minimal clinically important difference (MCID) 146 . For the HAM‐D, for example, a minimal clinically relevant improvement has been claimed by some authors to correspond to a 7‐point difference 147 or to an SMD of 0.88 148 . If this is correct, in psychotherapies or pharmacotherapies for depression, effect sizes of 0.30, 0.40 or even 0.50 correspond to a difference on the HAM‐D of 2 or 4 points (i.e. <7) which cannot be detected by clinicians and can therefore hardly be regarded as clinically significant. In schizophrenia, a SMD of 0.73 is required for a minimal clinical improvement of 15 points on the Positive and Negative Syndrome Scale (PANSS) 149 , implying that SMDs below 0.73 are not detectable by clinicians and may not be clinically significant.

For a better judgment, the CIs of the effect size estimates may be compared with the proposed MCID values 150 . It has been argued that, if the upper limit of the 95% CIs is smaller than the MCID, effect sizes can be regarded as “definitely clinically not important” 150 . For the vast majority of meta‐analyses on depression and schizophrenia, this would be the case if SMDs of 0.88 and 0.73 are used as MCID. However, even if the summary effect sizes are substantially smaller than the MCID, there is heterogeneity in treatment responses across patients. Therefore, a minority of patients may still achieve large benefits from treatment.

Fourth, identical effect sizes may have different clinical importance in different patient populations (e.g., according to disorders, gender or age) and for different outcomes (e.g., mortality) 151 . For outcomes including vital events (e.g., rates of suicide) small differences in success rates may be clinically important, whereas for continuous measures of (often transient) depression, anxiety or other symptoms, small differences of a few scale points may not 152 . Of the meta‐analyses on the treatment of depression, for example, only a few examined hard outcomes such as suicidal behaviour 41 , 42 , 44 . In the meta‐analyses on schizophrenia and bipolar disorder, data on suicidal behaviour were not reported, except for one meta‐analysis 118 . Future studies and meta‐analyses should include such important hard outcomes.

Fifth, TAU as a comparator was found to be superior to no treatment in depression (with regard to remission, 33% vs. 23%) 135 . However, TAU is a heterogeneous condition, and effect sizes achieved depend on the type of treatment actually delivered. Larger effect sizes may be achieved in comparison to weaker forms of TAU 23 , 153 . This applies to psychological placebo as well 154 .

Sixth, the results of RCTs may not necessarily represent real‐world effectiveness 155 . In clinical practice, patients often suffer from multiple mental disorders, and treatments are usually tailored to the individual patients' needs. This applies, for example, to treatment duration. Most of the treatments included in this review were short‐term 6 . Data on longer‐term treatments are widely missing from RCT research.

In summary, a systematic re‐assessment of recent evidence across multiple meta‐analyses on key mental disorders provided an overarching picture of limited additional gain for both psychotherapies and pharmacotherapies over placebo or TAU. A ceiling seems to have been reached with response rates ≤50% and most SMDs not exceeding 0.30‐0.40. Thus, after more than half a century of research, thousands of RCTs and millions of invested funds, the “trillion‐dollar brain drain” 2 associated with mental disorders is presently not sufficiently addressed by the available treatments. This should not be seen as a nihilistic or dismissive conclusion, since undoubtedly some patients do benefit from the available treatments. However, realistically facing the situation is a prerequisite for improvement. Pretending that everything is fine will not move the field forward 156 , nor will conforming and producing more similar findings 157 .

A paradigm shift in research seems to be required to achieve further progress. Suggestions for such a shift have recently been made 11 , for example, for improving methodological quality and replicability (e.g., open science 158 , 159 ), improving available treatments – e.g., by personalized management 160 , 161 , 162 , defining specific targets and outcomes 163 , considering response to previous treatments (staging) 164 , 165 , switching or augmentation strategies 166 – or developing new treatments (e.g., exploration of out‐of‐the box ideas and accidental discoveries 167 ). A focus on prevention (e.g., in educational or occupational settings) 168 , 169 may improve the situation as well.

Improving treatment strategies for mental disorders can be regarded as a central health challenge of the 21st century.

ACKNOWLEDGEMENTS

The authors are grateful to M. Huhn for providing effect sizes for the pharmacological treatment of schizophrenia. They thank J. Matzat, an official patient representative of the Federal Joint Committee in Germany, for approving the paper. They are also indebted to M. Wolf for his support in producing the figures, to S. Uysal for his help in the evaluation of the quality of the included meta‐analyses and the review of the extracted data, and to B. Leowald and L. Feix for support in searching for meta‐analyses. The paper is dedicated to the late H. Kächele, a pioneer of psychotherapy research. Supplementary information on the study is available at https://osf.io/yvg2c/ .

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Database of Abstracts of Reviews of Effects (DARE): Quality-assessed Reviews [Internet]. York (UK): Centre for Reviews and Dissemination (UK); 1995-.

Database of Abstracts of Reviews of Effects (DARE): Quality-assessed Reviews [Internet].

Comparing the effectiveness of individual therapy and group therapy in the treatment of depression: systematic review.

B Hodgkinson , D Evans , A O'Donnell , and K Walsh .

Review published: 1999 .

• Authors' objectives

To present the best available information on the use of group therapy and individual therapy in the treatment of long term depression. The primary question to be addressed was: for the treatment of long term depression, is group therapy or individual therapy appropriate and, if so, in what form? If neither is indicated, what modality might be better?

CINAHL, MEDLINE, PsycLIT, Current Contents, Science Citation Index, the Cochrane Library, DARE and EMBASE were searched (search terms and dates not provided). Bibliographies of all identified studies and review papers were also searched. Dissertation Abstracts International and proceedings databases were searched for unpublished studies.

• Study selection

Study designs of evaluations included in the review

Randomised or pseudo-randomised controlled trials (RCTs).

Specific interventions included in the review

Various forms of individual therapy (defined as any one to one interaction between the patient and the therapist) and group therapy (excluding family therapy). Studies involving only combined psychotherapy and pharmacotherapy and studies which involved pharmacotherapy alone were excluded, as were treatments combining both group and individual therapy.

Participants included in the review

Children or adults suffering from long term depression and a determined Beck Depression Inventory (BDI) value of 12 or above or a Hamilton Rating Score for Depression (HRSD) of 14 or above were examined. Studies having participants with accompanying psychiatric disorders (e.g. schizophrenia) were excluded.

Outcomes assessed in the review

Reduction in depression inventory scores such as BDI, HRSD.

How were decisions on the relevance of primary studies made?

Two reviewers assessed all identified abstracts for relevance. Studies identified from bibliography searches were assessed on study title.

• Assessment of study quality

Methodological quality was assessed using a checklist that was designed and trialed by the Joanna Briggs Institute for Evidence Based Nursing and Midwifery (JBIEBNM) and criteria included: randomisation, blinding, intention-to-treat analysis, comparability of groups at baseline, identical treatment of groups other than for named interventions, outcome assessment, statistical analysis. At least two reviewers were involved in assessing validity as the authors state that disagreements between reviewers were resolved by discussion with another reviewer.

• Data extraction

Data were extracted independently using a data extraction tool that was developed and tested prior to use by JBIEBNM. A separate reviewer dealt with disagreements.

• Methods of synthesis

How were the studies combined?

If appropriate with available data, results from comparable groups of studies were pooled by meta-analysis. Odds ratios (OR) and 95% confidence intervals (CI) were calculated for categorical data, and weighted mean differences (WMD) and 95% CIs for continuous data. Where possible intention-to-treat and/or completer analysis was performed. Where statistical pooling was not appropriate or possible, the findings were summarised in narrative form.

How were differences between studies investigated?

Heterogeneity between combined studies was tested using a standard chi-squared test.

• Results of the review

Nineteen RCTs and one systematic review. Numbers of participants were not clear.

Individual cognitive behavioural therapy versus pharmacotherapy (3 studies): both treatments were found to be effective in reducing BDI scores (n=155; WMD 1.3 favouring pharmacotherapy (95% CI: -1.3, 4.0)) and HRSD scores (n=138; WMD -0.6 favouring cognitive therapy (95% CI: -2.5, 1.2)). The authors note that the included studies predate selective serotonin reuptake inhibitors (SSRIs).

Individual cognitive therapy versus individual cognitive therapy combined with pharmacotherapy (2 studies): neither ICT nor combined therapy was found to be significantly more effective than the other in either trial. Due to the nature of the reporting of data, meta-analysis was not possible.

Individual cognitive therapy versus waiting list with medication support (1 study): ICT significantly reduced BDI scores over the period of treatment and at 6 month follow up compared to the control (p<0.05).

Cognitive individual and group therapy versus waiting list with support (treatment as usual) (2 studies): One study found significant reductions in BDI scores for group cognitive therapy compared to control at post treatment but the groups were not compared at follow-up. The other study found significant reductions in BDI scores for individual therapy compared to control at post treatment but not at 3 months follow-up. The two studies could not be combined in a meta-analysis.

Cognitive individual or group therapy versus waiting list (no other treatment) (2 studies): In one study group therapy generated significantly lower BDI and HRSD scores than control. In the other study, individual therapy was significantly more effective than control in reducing BDI and HRSD scores at post-treatment and at 2 month follow-up. Meta-analysis of these two studies could not be performed.

Cognitive group therapy versus individual cognitive therapy (4 studies): BDI scores post treatment (n=124) WMD 0.2 (95% CI: -2.1, 2.6) and HRSD scores (2 studies, n=42)) WMD -0.3 (95% CI: -3.6, 3.0). Follow up BDI scores were not significantly different between groups at 2 and 3 months, but at 6 months CGT was favoured (n=62; WMD -6.9, 95% CI -11.6, -2.2). One study using ITT analysis claimed a significant advantage of ICT over CGT in BDI scores at post treatment and follow-up (p<0.02) and two did not.

Individual psychotherapy versus individual cognitive therapy (1 study): no significant difference between treatments was shown.

Cognitive group therapy versus computer assisted therapy (therapeutic learning programme or TLP) (2 studies): both studies reported no significant difference in the effectiveness of treatments in reducing depression scores; meta-analysis could not be performed.

Coping with depression: a course, comparison of group or class therapy and individual therapy (2 studies): Meta-analysis showed individual treatment was significantly more effective at reducing BDI scores than was the class method (n=104; WMD 3.0, 95% CI 1.0, 5.0), however the effect did not persist at follow up at 1 and 6 months.

Individual cognitive therapy versus waiting list control (1 systematic review): Remission from depressive disorder was higher in the ICT group compared to control OR 2.2 (95% CI: 1.4, 3.5).

Cognitive group therapy versus waiting list control (2 studies): CGT was significantly better than control in reducing BDI scores post-treatment (n=56; WMD -11.2, 95% CI -16, -6.5). One study found this difference persisted at follow-up. CGT was also significantly better than control for producing participants with normal BDI scores post-treatment (n=46; Peto OR 11.8, 95% CI 3.3, 42.3).

• Authors' conclusions

Individual and group cognitive behavioural therapies for moderately or severely depressed adults (BDI 14 or above) are comparable with each other in effectiveness and both are superior to providing no treatment at all. Individual cognitive therapy is equal to or better than tricyclic antidepressant drugs given at recommended therapeutic dosages for depressed people with a mean BDI of 30. This information is based on level II evidence (RCT).

• CRD commentary

This is a good review. The inclusion criteria are clearly defined relative to the research question and the literature search is very comprehensive so it is unlikely that any studies would have been missed, although it is not stated whether studies were restricted by language. Details of the review process are given and validity was assessed but not used. Study details were not fully presented and could have been presented in an appendix. It may have been more appropriate to use relative risk rather than odds ratio as a summary estimate. Each comparison is based on quite a small number of participants in a few trials, and some reported only pre- versus post-treatment data rather than results of a randomised comparison. The authors' conclusions do follow from the results but should be treated with some caution owing to flaws in methodological quality of the included trials reported in the text.

• Implications of the review for practice and research

Practice: The authors state practice implications for adults and adolescents.

For adults, either group (CGT) or individual cognitive behavioural therapy (ICT) can be used to treat moderate to severely depressed patients with the choice of therapy dependent upon the clinicians perceived receptiveness of the individual patient to group versus individual treatment. The use of computer assisted therapy can be useful as an aid to CGT in moderate to severely depressed patients. ICT can be effective in place of pharmacotherapy in moderate to severely depressed patients if the patient is opposed to being treated with drug therapy. CGT has not been compared to pharmacotherapy so no direct recommendation can be given as to its effectiveness as a replacement therapy.

For adolescents, either group or individual cognitive behavioural therapy can be used to treat moderately depressed adolescents (BDI 14 or above).

Research: The authors state that more research is needed to determine the effectiveness of individual or group cognitive behavioural therapy in severely depressed adolescents (BDI 20 or above).

• Bibliographic details

Hodgkinson B, Evans D, O'Donnell A, Walsh K. Comparing the effectiveness of individual therapy and group therapy in the treatment of depression: systematic review. Adelaide, S. Australia, Australia: Joanna Briggs Institute for Evidence Based Nursing and Midwifery. Systematic Review; 3. 1999.

• Indexing Status

Subject indexing assigned by CRD

Cognitive Therapy; Depressive Disorder /therapy

• AccessionNumber

11999009777

• Database entry date
• Record Status

This is a critical abstract of a systematic review that meets the criteria for inclusion on DARE. Each critical abstract contains a brief summary of the review methods, results and conclusions followed by a detailed critical assessment on the reliability of the review and the conclusions drawn.

• Cite this Page Hodgkinson B, Evans D, O'Donnell A, et al. Comparing the effectiveness of individual therapy and group therapy in the treatment of depression: systematic review. 1999. In: Database of Abstracts of Reviews of Effects (DARE): Quality-assessed Reviews [Internet]. York (UK): Centre for Reviews and Dissemination (UK); 1995-.

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The Effectiveness of Psychological Interventions Delivered in Routine Practice: Systematic Review and Meta-analysis

• Original Article
• Open access
• Published: 06 October 2022
• Volume 50 , pages 43–57, ( 2023 )

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• Chris Gaskell   ORCID: orcid.org/0000-0002-7589-5246 1 ,
• Melanie Simmonds-Buckley   ORCID: orcid.org/0000-0003-3808-4134 1 ,
• Stephen Kellett   ORCID: orcid.org/0000-0001-6034-4495 1 , 2 ,
• C. Stockton 1 ,
• Erin Somerville 1 ,
• Emily Rogerson 1 &
• Jaime Delgadillo   ORCID: orcid.org/0000-0001-5349-230X 1  

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This review presents a comprehensive evaluation of the effectiveness of routinely delivered psychological therapies across inpatient, outpatient and University-based clinics. This was a pre-registered systematic-review of studies meeting pre-specified inclusion criteria (CRD42020175235). Eligible studies were searched in three databases: MEDLINE, CINAHL and PsycInfo. Pre–post treatment (uncontrolled) effect sizes were calculated and pooled using random effects meta-analysis to generate effectiveness benchmarks. Moderator analyses were used to examine sources of heterogeneity in effect sizes. Overall, 252 studies ( k  = 298 samples) were identified, of which 223 ( k  = 263 samples) provided sufficient data for inclusion in meta-analysis. Results showed large pre–post treatment effects for depression [ d  = 0.96, (CI 0.88–1.04), p  ≤ 0.001, k  = 122], anxiety [ d  = 0.8 (CI 0.71–0.9), p  ≤ 0.001, k  = 69], and other outcomes [ d  = 1.01 (CI 0.93–1.09), p  ≤ 0.001, k  = 158]. This review provides support for the effectiveness of routinely delivered psychological therapy. Effectiveness benchmarks are supplied to support service evaluations across multiple settings.

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Introduction

Meta-analyses of clinical trials support the efficacy of psychological interventions for various mental health problems such as depression (Cuijpers et al., 2008 ), anxiety disorders (e.g., Cuijpers et al., 2014a ; Mayo-Wilson et al., 2014 ; Olatunji et al., 2014 ; Sánchez-Meca et al., 2010 ; Wolitzky-Taylor et al., 2008 ), post-traumatic stress disorder (Lewis et al., 2020 ), obsessive–compulsive disorder (Rosa-Alcázar et al., 2008 ), eating disorders (Linardon et al., 2017 ) and other conditions. Grounded in this evidence, clinical guidelines support the use of psychological interventions in routine clinical care (e.g., Chambless & Hollon, 1998 ; Chambless & Ollendick, 2001 ; National Institute for Health & Care Excellence, 2011 ). These guidelines commonly advocate the implementation of empirically supported treatments, closely following the procedures implemented in clinical trials and specified in associated treatment manuals. To this end, competency frameworks have been developed to support the dissemination of empirically supported treatments in routine care and clinical training programmes (e.g., Lemma et al., 2008 ; Roth & Pilling, 2008 ; Roth et al., 2009 ).

Some studies have found similar treatment outcomes when comparing data from efficacy trials and routine practice (e.g., Lutz et al., 2016 ; Persons et al., 1999 ). However, there are some reasons to assume that the effects of psychotherapy delivered in routine care settings may differ from those observed in clinical trials. Recent evidence indicates that psychological treatment outcomes are associated with treatment integrity , which refers to the competent (skilled) delivery of protocol-driven treatment procedures (Power et al., 2022 ). However, surveys of clinicians working in routine settings often reveal negative attitudes towards protocol-driven treatment and a lack of adherence to treatment manuals (e.g., Addis & Krasnow, 2000 ). Hence, the integrity of routinely delivered psychological treatments is unclear, and it probably varies across services (Freedland et al., 2011 ). Furthermore, the strict selection criteria applied in clinical trials may result in unusually homogeneous samples that do not reflect the diverse clinical populations typical of routine care settings (e.g., Lambert, 2013 ; Zimmerman et al., 2002 ). Previous studies have found systematic differences in the clinical profiles of patients included and excluded from psychotherapy trials (e.g., van der Lem et al., 2012 ). For these reasons, it is plausible to assume that the effects of routinely delivered therapy may vary across settings and clinical populations, and may not necessarily conform to benchmarks from efficacy trials.

A tradition of practice-based evidence (PBE, Margison et al., 2000 ) has emerged in recent decades, with numerous studies examining the effects of routinely delivered psychological interventions in various settings. Narrative reviews of PBE generally confirm that moderate-to-large uncontrolled (pre-to-post treatment) effect sizes are observed in routine care settings, supporting the effectiveness of psychotherapy but also demonstrating considerable variability across patient samples, therapists and clinics (e.g., see Barkham et al., 2010 ; Castonguay et al., 2013 , 2021 ). An inherent limitation of such narrative reviews is that they perform a selective rather than systematic synthesis of available data. Benchmarking studies can be useful to provide general indices of treatment effectiveness, enabling services to evaluate their outcomes relative to efficacy trials (e.g., McAleavey et al., 2019 ; Minami et al., 2008 ) or aggregated effect size data from similar clinical services (e.g., Delgadillo et al., 2014 ). Psychotherapy benchmarking studies tend to report favorable pooled effects sizes, but also show variability in effects across clinics (e.g., Barkham et al., 2001 ; Connell et al., 2007 ; Delgadillo et al., 2014 ; Gyani et al., 2013 ). Although benchmarking studies help to quantify the expected magnitude of treatment effects observed in routine clinical settings, most are nevertheless circumscribed to small sets of clinics or geographical areas, offering limited insights into possible sources of heterogeneity in treatment outcomes. Systematic reviews and meta-analyses may therefore offer a more comprehensive examination of the effectiveness of routinely delivered treatments.

Some meta-analytic investigations have reported that outcomes from routine practice-based treatments are not as favorable as those delivered in research settings (Weisz et al., 1995 ). Other meta-analyses suggest that there are no differences in treatment effects when comparing PBE and efficacy studies after controlling for case-mix differences (e.g., Shadish et al., 1997 , 2000 ). However, many of the PBE studies in these meta-analyses applied stringent controls on the treatment procedures (e.g., adherence and competence assessments)—making them more akin to efficacy trials. Hunsley and Lee ( 2007 ) reviewed 35 studies and concluded that the completion and improvement rates observed in PBE studies were comparable to efficacy trials. Cahill et al. ( 2010 ) reviewed 31 studies, concluding that psychotherapy was most effective for the treatment of common mental disorders, with a pooled uncontrolled effect size of d  = 1.29. More recently, Wakefield et al. ( 2021 ) reviewed 60 studies, of which 47 were eligible for meta-analysis. They reported large uncontrolled effect sizes for depression ( d  = 0.87) and anxiety ( d  = 0.88), and a moderate effect on functional impairment ( d  = 0.55). These meta-analyses show wide variability in treatment effects (i.e., heterogeneity) across studies/samples.

PBE meta-analyses provide some insights into plausible sources of heterogeneity, including methodological (e.g., completers analyses vs. inclusion of patients lost to follow-up) and clinical features (e.g., larger effects for common mental disorders, lower effects for patients with comorbidities and socioeconomic disadvantages, larger effects for lengthier interventions). Nevertheless, these meta-analyses are over a decade old (Cahill et al., 2010 ; Hunsley & Lee, 2007 ) or limited to a specific treatment setting (e.g., primary care outpatient services; Wakefield et al., 2021 ). Further research into the methodological and clinical sources of treatment heterogeneity is needed to better understand why treatment effects vary across samples, and to determine whether or not these effects vary across different treatment settings (e.g., outpatient, inpatient, university-based treatment).

The considerable growth of the PBE literature in the last decade and Implementation of empirically supported treatments across many settings warrants a comprehensive review of treatment outcomes data. The aim of the present study was to systematically review available PBE studies. The objectives of the study were to (1) provide benchmarks of treatment effectiveness using meta-analysis and (2) to examine sources of effect size heterogeneity using pre-specified moderator analyses informed by earlier studies.

Search Strategy and Study Selection

The present study followed good practice guidelines for systematic reviews (PRISMA, Page et al., 2021 ) and meta-analyses of psychotherapy studies (MAP-24, Flückiger et al., 2018 ). A review protocol was pre-registered in the PROSPERO database (CRD42020175235). Footnote 1 Literature searches were carried out without any restrictions on date of publication up to the search date (April 2020). Inclusion criteria were: (a) studies reporting outcomes for routinely delivered treatments (i.e., not as part of efficacy trials); (b) all adult sample (no patients under 16); (c) employed a psychological treatment (i.e., driven by psychological theory and intended to be therapeutic (Spielmans & Flückiger, 2018 ), as inferred or described by study manuscripts); and (d) conducted face-to-face. Studies were excluded if they: used (e) family/group treatments, (f) were not available in English; (g) did not employ a self-report measure of treatment effectiveness Footnote 2 ; (h) did not provide sufficient data to calculate pre–post treatment effect sizes; or (i) employed randomization procedures or control groups. A more detailed table of inclusion/exclusion criteria is available in supplementary Table 1.

The search strategy had three phases. Phase one was a systematic search of three electronic literature databases (MEDLINE, CINAHL and PsycInfo) via EBSCO using a pre-registered list of key terms. Methodological terms included: practice-based evidence , routine practice , benchmarking , transportability , transferability , clinically representative , managed care setting , uncontrolled , external validity , applicable findings , empirically supported , dissemination , and clinical effectiveness evaluation . These terms were informed by prior reviews of psychotherapy effectiveness (Cahill et al., 2010 ; Stewart & Chambless, 2009 ). Effectiveness and evaluation were not used as single word terms due to producing unmanageable numbers of irrelevant records. For the psychologically relevant term: psycho * OR therap * was used for PsycInfo while psycho * alone was used for MEDLINE and CINAHL ( therap * was removed from MEDLINE/CINAHL due to producing an unmanageable number of irrelevant records). Limiters included adult population and English language . No exclusions were made based on the type of publication. Key term combinations and Boolean operators are reported in supplementary Table 2. Phase two included a manual search of reference lists, and forward citation searching (using Google Scholar) for studies identified in phase one. Titles relevant to the current review were identified by the first author. Finally, phase three was a grey literature search using the terms psychotherapy AND routine-practice AND effectiveness in Google Scholar.

After removal of duplicates, titles and abstracts of potentially eligible studies were screened by the first author using a pre-developed and piloted screening tool. Sub-samples were screened by a second coder at each stage (20% at the stage of title screening; 10% at the stage of full-text screening). Percentage agreement and inter-rater reliability statistics (Kappa ( \(\kappa\) ), Cohen, 1960 ] indicated good reliability ( \(\kappa\) = 0.78, 1713/1740, 98.45%) in the first stage and adequate reliability ( \(\kappa\) = 0.65, 24/30, 80%) in the second stage. After the selection process was completed, corresponding authors for eligible studies were contacted via email to request additional recommendations for potentially eligible studies, and to request additional statistical information to calculate effect sizes. E-mail responses were received from 76 authors and additional data was provided for 41 samples.

Data Extraction

There were three separate outcome domains (and subsequently three meta-analyses) for ‘depression’, ‘anxiety’ and ‘other’ outcomes. The latter category consisted of general psychological distress scales, measures of functioning/quality of life, or diagnosis-specific outcome scales (e.g., obsessive-compulsive disorder, etc.). A pilot extraction sheet was developed and pilot-tested with a sample of studies ( k  = 10). When multiple samples were reported in the same study, effect-sizes across these samples were aggregated to reduce bias of statistical dependency (Gleser & Olkin, 2009 ; Hoyt & Del Re, 2018 ). To avoid loss of information (e.g., aggregating sub-samples that are distinct based on levels of a moderator), study samples were disaggregated for moderator analyses (Cooper, 1998 ). Studies with overlapping datasets (e.g., reanalysis of the same sample) were only included once in the meta-analysis. Samples which performed an intention-to-treat (ITT) analysis were preferred to completer samples due to being less prone to attrition bias (Jüni et al., 2001 ); so the ITT data was extracted for studies that reported both ITT and completer analyses. As extraction of multiple study effect-sizes within a single domain (e.g., depression) threatens statistical dependency (Borenstein et al., 2021 ) we selected a single effect-size per domain (Card, 2015 ; Cuijpers, 2016 ), using a preference system (defined a priori, supplementary material). Reliability of coding for effect-size data was computed using a second coder for a sub-sample of manuscripts (n = 29) demonstrating almost perfect reliability across all values ( \(\kappa\) = 0.97, agreement = 97.56%) and perfect reliability for effect-size values ( \(\kappa\) = 1.00). Key categorical and numerical variables extracted from manuscripts for moderator analyses are reported in Table 1 . For sample severity, the decision was made to cluster university counselling centers in the ‘mild’ severity category due to prior research finding normative data of UK University students comparable to primary care samples (Connell et al., 2007 ).

Risk of Bias and Quality Assessment

The Joanna Briggs Institute Quality Appraisal Tool for Case Series (Munn et al., 2020 ) was used to assess risk of bias. Eight criteria primarily focusing upon manuscript reporting detail were used. Criteria included manuscript reporting of: (i) patient inclusion criteria, (ii) service description, (iii) treatment description, (iv) sample characteristics, (v) outcome data, (vi) effect-size calculation, (vii) consecutive patient recruitment, and (viii) inclusion of patients lost to follow-up in statistical analysis. Each item was coded as either met or not met (including not clear) by the first author for each sample. A sub-sample (23.8%) was rated independently by two other reviewers (11.9% each). The pooled agreement was 84.17% ( \(\kappa\) = 0.62).

Statistical Analysis

All analyses were conducted using the R statistical analysis environment (R Core Team, 2020 , v 4.0.2). We calculated standardised mean change (SMC: Becker, 1988 ) for included studies using the metafor package. This approach divides the pre–post mean change score by the pretreatment standard deviation with a sampling variance adjustment using the correlation between the pre-treatment and post-treatment measures (Morris, 2008 ). When unavailable, Pearson’s r was imputed using an empirically derived estimate ( r  = .60, Balk et al., 2012 ). Aggregation of samples/sampling errors was conducted using the aggregate function of metafor using standard inverse-variance weighting.

Random effects meta-analyses were performed using the metafor (Viechtbauer, 2020 ), dmetar (Harrer et al., 2019 ), and meta (Schwarzer, 2020 ) packages. Forest plots were used to visualise pre–post treatment effects sizes across samples. Effect size heterogeneity was assessed using I 2 (Higgins & Thompson, 2002 ) and the Q statistic (Cochran, 1954 ). Publication bias was examined using funnel plots and assessed statistically using rank correlation tests (Begg & Mazumdar, 1994 ), Egger’s regression test for funnel plot asymmetry (Egger et al., 1997 ), and the fail-safe N (Rosenthal method, Rosenthal, 1979 ).

Moderator analyses were based on a set of moderator variables selected a priori, following evidence from prior reviews. Subgroup variables included: (i) analysis (inclusion of patients lost to follow-up), (ii) geographical region , (iii) severity (mild, moderate, severe, university Footnote 3 ), (iv) treatment modality , (v) experience [unqualified (i.e., trainees) vs. qualified therapists], (vi) stage of treatment development (preliminary study vs. routine evaluations), and (vii) sample size (small, medium, large). Continuous meta-regression variables included (i) publication year , (ii) average age of sample, and (iii) percentage of samples who identified as female . All moderators were included in meta-regression which was based on a mixed effects (i.e., multilevel) model (Borenstein et al., 2021 ) with weighted estimation (inverse-variance weights).

Finally, we developed effect size benchmarks to support the evaluation of effectiveness across four broad settings: outpatient services, inpatient services and university counselling services (i.e., student population) and university psychotherapy clinics (non-student population). Informed by previous benchmarking studies (Delgadillo et al., 2014 ), pooled effect sizes (using random effects meta-analyses) were stratified into quartiles to differentiate between low effectiveness (bottom 25%), average effectiveness (middle 50%) and high effectiveness benchmarks (top 25%).

Search Results

The PRISMA diagram in Fig.  1 presents a summary of the study selection process. Overall, 10,503 records were identified, of which 252 manuscripts were eligible for inclusion and 223 (samples k  = 263) had sufficient information to be included in the meta-analysis. Summary statistics are provided in Table 2 .

Prisma flow diagram of studies throughout the review

Study Characteristics

Eligible studies were published between 1984 and 2020 (median = 2013, k  = 294 published ≥ 2000). Of these, 169 samples included patients lost to follow-up ( k  = 118, 56.72% completers). Most studies were from the USA ( k  = 113, 37.92%), England ( k  = 78, 26.17%), Germany ( k  = 24, 8.05%), Sweden ( k  = 12, 4.02%) and Canada ( k  = 10, 3.36%). These five most represented countries accounted for most of the included samples ( k  = 237, 79.53%).

Sample Characteristics

Sample characteristics were reported for 291 samples, with a cumulative N of 233,140 patients (mean = 838.63, median = 81.5, range—4 to 33,243, IQR = 224.5). The prevalence of female participants was 61.88% (N = 144,273, k  = 279) with 13 all-female samples and 2 all-male samples. The mean average sample age was 35.33 years (range = 19.00–60.50). Across studies which provided information, 23.00% of patients were from ethnic minorities ( k  = 127), 37.00% were married ( k  = 106), and 23.00% were in employment ( k  = 96).

Treatment Characteristics

Most samples evaluated cognitive-behavioral interventions ( k  = 152, 51.01%) while 50 samples evaluated psychodynamic (16.78%), and 25 samples evaluated counselling (8.29%; other = 71, 23.82%). Counselling interventions were interventions described simply as ‘counselling’ by study authors (with no further treatment information) or ‘person-centered counselling’ interventions. Interventions termed ‘counselling’ but described in a way that fit closely with of the other treatment modalities (e.g., cognitive-behavioral counselling) was assigned to the more specific treatment modality group. For symptom severity, 96 (32.21%) samples came from services treating mild conditions, 92 (30.87%) from services treating moderate conditions, 33 (11.07%) from services treating severe conditions, and 68 (22.82%) from university psychotherapy clinics (not counselling centers) that treated a wide spectrum of conditions from mild-to-severe (other, k  = 9, 3.02%). Treatment dosage, when reported ( k  = 256) was in hours/sessions ( k  = 225), months ( k  = 12) or days ( k  = 8). The pooled (non-weighted) average dose (hours) was 16.30 sessions (median = 13.00, range = 1.00–139.30, IQR = 11.00). A total of 62 (20.81%) samples reported that treatment was delivered exclusively by trainees, while 100 (35.58%) samples reported having at least one trainee.

Risk of Bias

In order of satisfactory criteria (e.g., the criterion under evaluation was met), the following risk of bias domains were assessed: demographic reporting detail (264/298, agreement = 98.33%, \(\kappa\) = 0.88), service reporting detail (260/298, agreement = 85%, \(\kappa\) = 0.31), study outcome reporting details (240/298, agreement = 83.33%, \(\kappa\) = − 0.03), intervention reporting detail (234/298, agreement = 85%, \(\kappa\) = 0.32), service inclusion criteria (214/298, agreement = 90%, \(\kappa\) = 0.64), appropriate use of analysis (214/298, agreement = 70%, \(\kappa\) = 0.26), complete inclusion (i.e. consecutive recruitment and inclusion of those lost to follow-up, 41/298, agreement = 85%, \(\kappa\) = 0.45), and consecutive inclusion (93/298, agreement = 76.67%, \(\kappa\) = 0.51).

Meta-analyses

The random-effects meta-analysis for depression outcomes ( k  = 140, N = 68,077), across 10 unique measurement tools was statistically significant ( p  ≤ 0.001), indicative of a large pre–post treatment ( d  = 0.96, CI 0.88–1.04) reduction in depression severity. There was a large magnitude of statistically significant heterogeneity [I 2  = 97.94%, Q(df = 121) = 2677.37, p  ≤ 0.001]. The funnel plot (Fig.  2 ) shows limited visual evidence of asymmetry. The funnel rank correlation test was not statistically significant ( \(\tau\) = 0.061, p  = 0.46) however the funnel regression test was statistically significant (Z = 2.13, p  = 0.033). The fail-safe N was 515,853.

Funnel plots displaying the distribution of studies reporting pre–post outcomes for (i) depression, (ii) anxiety, and (iii) miscellaneous outcomes

The random-effects meta-analysis for anxiety outcomes ( k  = 84, N = 26,689, measurement tools = 20) was statistically significant ( p  ≤ 0.001), indicative of a large ( d  = 0.80, CI 0.71–0.90) reduction in symptom severity. Heterogeneity was large and statistically significant [I 2  = 97.51%, Q(df = 68) = 1328.96, p  ≤ 0.001]. The funnel plot shows limited evidence of asymmetry. The funnel rank correlation test was not significant ( \(\tau\) = 0.009, p  = 0.888). In contrast, the funnel regression test was statistically significant (Z = 2.533, p  = 0.011). The fail-safe N was 121,899.

The random-effects meta-analysis for other outcomes ( k  = 184, N = 126,734, measurement tools = 40) was statistically significant ( p  ≤ 0.001), indicative of a large ( d  = 1.01, CI 0.93–1.09) reduction in severity of indices of distress. Heterogeneity was large and statistically significant [I 2  = 99.06%, Q(df = 157) = 15,330.32, p  ≤ 0.001]. The funnel plot shows a degree of asymmetry with clustering to the right of the mid-line. The funnel rank correlation test was statistically significant ( \(\tau\) = 0.208, p  ≤ 0.001). In contrast, the funnel regression test was not significant (Z = 3.697, p  ≤ 0.001). The fail-safe N was 1,695,607.

Moderator Analyses

Multivariable meta-regressions were conducted for each of the three outcome domains (Tables 3 , 4 , 5 ). After controlling for other moderators, the depression meta-regression found a significant effect for geographical region, therapist experience and type of analysis. UK samples had larger effect sizes compared to samples from Asia; effects sizes in samples treated by qualified staff members were larger than those observed in samples exclusively consisting of trainees; and samples excluding patients lost to follow-up (i.e., completer analyses) had larger effect sizes compared to intention-to-treat analyses. For anxiety outcomes, UK studies had larger effect sizes than studies from mainland Europe; mild severity samples had larger effect sizes than samples of patients with moderate or severe symptoms; and cognitive-behavioural interventions had larger effect sizes than counselling interventions. Finally, for other outcomes, the only significant moderator indicated that cognitive-behavioural interventions had larger effect sizes than psychodynamic interventions and unspecified (i.e., other) interventions.

Benchmarking Data

Pooled effect-sizes for low, average and high performing services are shown in Table 6 , organized according to setting [outpatient services, inpatient services, university counselling services (i.e., student population) and university psychotherapy clinics (non-student population)]. Although the effect size estimates for each benchmark vary across settings, confidence intervals consistently overlapped, indicating similar levels of symptom-changes across the performance strata (low, average, high). The exception to this is the low performance benchmark for anxiety measures which were significantly larger in university psychotherapy clinics ( d  = 0.51) and significantly smaller in inpatient services ( d  = 0.13) by comparison to outpatient services ( d  = 0.37).

This review provides a comprehensive quantitative review of the effectiveness of psychological treatments delivered in routine care settings. Overall, 252 studies (samples k  = 298) were identified, of which 223 (88.5%, k  = 263) were included in the meta-analysis. Consistent with prior psychotherapy effectiveness reviews, we found large uncontrolled (pre–post treatment) effect sizes ( d  = 0.80–1.01) across multiple outcome domains (depression, anxiety, and general psychological distress).

Consistent with previous meta-analyses of PBE (e.g., Cahill et al., 2010 ; Hunsley & Lee, 2007 ; Wakefield et al., 2021 ), we observed wide variability in effect sizes across studies and large (> 90%) indices of heterogeneity across outcome domains. The large number of samples included in this review enabled us to carry out adequately-powered moderator analyses to better understand potential sources of heterogeneity. For depression outcomes, smaller effect sizes were found for samples in Asia (compared to the UK), and in treatments delivered by trainees (i.e., compared to qualified professionals). For anxiety outcomes, smaller effect sizes were found for treatments delivered in mainland Europe (compared to the UK), services treating patients with moderate or high levels of severity (compared to mild severity), and counselling interventions (compared to cognitive-behavioural interventions). For other outcomes, only therapy modality was significant. Psychodynamic and unspecified interventions produced smaller effect-sizes (compared to cognitive-behavioural interventions). To some extent, these results are consistent with and support clinical guidelines that recommend cognitive-behavioural therapy as a first-line intervention, prior to considering other treatment modalities (National Institute for Health & care Excellence, 2011 ). However, caution is advised when interpreting these between-therapy comparisons using uncontrolled data from observational studies, as they could be explained by other unmeasured factors such as relevant case-mix differences between patients (e.g., socioeconomic status, personality, comorbid physical illnesses, etc.). Studies that control for case-mix variables using individual patient data find that there are no significant differences in treatment effects when comparing different treatment modalities (e.g., Pybis et al., 2017 ). Furthermore, as found in a previous meta-analysis (Wakefield et al., 2021 ), completers analyses tended to produce inflated (biased) effect sizes by comparison to intention-to-treat (more conservative and stringent) analyses.

The finding of large clinical improvements during psychotherapy and across outcomes was consistent with prior meta-analyses of psychotherapy effectiveness for depression outcomes (Hans & Hiller, 2013 ; Wakefield et al., 2021 ), anxiety outcomes (Stewart & Chambless, 2009 ; Wakefield et al., 2021 ), and other indices of psychological distress and functioning (Cahill et al., 2010 ). Pooled uncontrolled effect-sizes were smaller than that reported by Cahill et al. ( 2010 ) ( d  = 1.29), although this may reflect differences in the focus of the reviews (e.g., Cahill et al., 2010 included group treatments) or the changing distribution of geographical representation (i.e., more studies from non-UK/North American countries). Large clinical improvements are also consistent with many meta-analyses of psychotherapy controlled trials (e.g., Cuijpers et al., 2008 , 2014a ; Mayo-Wilson et al., 2014 ; Olatunji et al., 2014 ).

It is possible that there are continental differences in models of training, service structures, therapy provision and emphasis on evidence-based practice which underlie the observed differences in pooled effect-sizes between continents. This is consistent with UK and US clinical guidance recommending delivery of empirically supported treatments (APA, 2006 ; National Institute for Health and Care Excellence, 2011 ). It is possible that the service policy context in the UK places greater emphasis on the delivery of treatment with high fidelity to empirically supported treatment protocols, and this may explain the relatively larger effect sizes in this geographical location, since high integrity is associated with better treatment outcomes and especially for anxiety treatment outcomes (Power et al., 2022 ). Despite these differences, all continents demonstrated positive change for all outcomes ( d  = 0.59–1.10) supporting the universality hypothesis (i.e., that psychotherapy is assumed to work across cultures; Flückiger et al., 2018 ).

Consistent with several prior meta-analytic reviews (e.g., Cuijpers et al., 2014b ; Driessen et al., 2010 ; Furukawa et al., 2017 ), symptom severity did not predict effectiveness of treatment for depression. For anxiety outcomes, services categorized as treating mild conditions consistently had larger effect sizes. It is possible that classifying by type of service provided an imprecise proxy for sample severity and therefore future research should explore severity as a continuous variable in routine settings.

Limitations

The most notable critique of this review is that it is based exclusively on evidence from observational studies. We are unable to rule out alternative explanations for observed effect sizes [placebo effects, spontaneous remission (Posternak & Miller, 2001 ; Whiteford et al., 2012 )] and subsequently the observed effect sizes in this review cannot be directly compared to efficacy trials. Nevertheless, pooled effect sizes from observational studies serve as a valuable data source for benchmarking of routine care and quality improvement initiatives (e.g., Clark et al., 2018 ; Delgadillo et al., 2014 ; Gyani et al., 2013 ).

A key design limitation concerns statistical dependency. Efforts to avoid statistical dependency included: (i) taking one sample measure per domain, (ii) aggregating multiple unique study samples within a single domain, and (iii) extracting one measurement tool per study, per construct (i.e., preference system). These approaches have well-documented limitations (Borenstein et al., 2021 ; Hoyt & Del Re, 2018 ; Van den Noortgate et al., 2013 ). A preferable approach would have been to model dependency using a multi-level analysis (Van den Noortgate et al., 2013 , 2015 ) or through robust variance estimation and should be considered for future replications. Use of robust-variance estimation would avoid the need to assign outcomes to a restrictive number of outcome domains. This would also circumvent the need to adopt a highly heterogeneous “other” outcome domain, which for the current review included both diagnosis specific and global distress-based measures.

An additional limitation concerns the inherent limitations of the risk-of-bias assessment tool which was selected for this study a priori. It could be argued that this tool primarily indexes manuscript reporting detail and not necessarily risk of bias. Future reviews of effectiveness could consider assessing methodological rigour using other available rating tools (e.g., see Munder & Barth, 2018 ).

Due to resource constraints and the large number of included studies, the systematic search, data extraction and risk-of-bias ratings were not performed completely in duplicate. For the subsample of full texts screened by two coders there was a strong, but imperfect, agreement/reliability (80%, \(\kappa\) = 0.65). Similarly, not extracting data or assessing RoB in duplicate is problematic due to risk of imprecise estimates of treatment effect and RoB (Armijo-Olivo et al., 2014 ). An additional limitation surrounds coding decisions for moderator variables. Therapy modality was coded from manuscript self-definition. The degree to which treatments truly resembled treatment code (or treatment intended) is not clear. It was also apparent during extraction that very few practice-based studies report fidelity/adherence checks. As this becomes more routinely reported opportunities for modelling differences based on adherence/competence/integrity will become available. The use of categorical moderator levels to differentiate samples at the study level may also have provided imprecise proxies for moderator levels. For example, patient severity would preferably be modelled through meta-regression at the patient level to account for the heterogeneity within samples as it has been shown that university counselling center samples have numerous highly distressed individuals (Xiao et al., 2017 ). Future studies investigating these moderator variables at the patient level (e.g., through individual participant data meta-analysis) would help to shed light on this.

The search strategy is unlikely to have identified every available study. Search terms were based on prior reviews and omitted several terms that were found to produce an unmanageable number of records (e.g., “effectiveness”, “evaluation”). Despite this, the current reviews gives an adequate range and depth of effectiveness research with which to make tentative interpretations regarding the field of psychotherapy effectiveness research. A final caveat is the decision to focus exclusively on self-report measures of effectiveness. Meta-analytic evidence has demonstrated significant differences between self-report and clinician rated measures of clinical improvement (Cuijpers et al., 2010 ). Future research is therefore needed to see if the pooled effect-sizes from this study are consistent with clinician-rated measures of effectiveness in routine settings.

Conclusions

This review provides support for the effectiveness of psychological therapy as delivered in routine settings across a range of outcomes. Overall, the effects of psychotherapy appear to generalize well to diverse clinical settings, contexts, and populations. Nevertheless, it is evident that treatment effects vary considerably across services, and this review provides performance benchmarks to support routine service evaluation and practice development initiatives.

Data Availability

Data for the systematic review and related code will all be made publicly available through the lead author’s GitHub account following publication.

Standards of Reporting

This review followed the PRISMA and MAP-24 reporting guidelines for conducting systematic review/meta-analysis.

Protocol available at: https://www.crd.york.ac.uk/prospero/display_record.php?RecordID=175235 .

The authors recognise that use of the term effectiveness may be somewhat misleading. The pre–post (uncontrolled) methodology which forms the body of evidence in this review is unable to disentangle treatments effects from other potential causes of change (e.g., regression to the mean, placebo). Observed change in symptoms may therefore not exclusively represent treatment effectiveness. We have opted to retain use of this term within the current review because it has consistently and frequently been used as such in the extant literature (e.g., Lambert, 2013 ; Nordmo et al., 2020 ).

University clinics refers to university managed clinics treating communities beyond the student population. University counselling centres that are more specifically targeted at the student population are included within the mild category.

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Gaskell, C., Simmonds-Buckley, M., Kellett, S. et al. The Effectiveness of Psychological Interventions Delivered in Routine Practice: Systematic Review and Meta-analysis. Adm Policy Ment Health 50 , 43–57 (2023). https://doi.org/10.1007/s10488-022-01225-y

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DOI : https://doi.org/10.1007/s10488-022-01225-y

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