Clinical Psychology and Psychotherapy Clin. Psychol. Psychother. 23, 217–225 (2016) Published online 15 March 2015 in Wiley Online Library (wileyonlinelibrary.com). DOI: 10.1002/cpp.1948
The German Version of the Behavioral Activation for Depression Scale (BADS): A Psychometric and Clinical Investigation Tobias Teismann,1* Andrea Ertle,2 Nadine Furka,3 Ulrike Willutzki4 and Juergen Hoyer5 1
Department of Clinical Psychology and Psychotherapy, Ruhr-Universität Bochum, Bochum, Germany Institute of Psychology, Humboldt-Universität zu Berlin, Berlin, Germany 3 Institute of Clinical Psychology and Psychotherapy, Technische Universität Dresden, Dresden, Germany 4 Department for Psychology and Psychotherapy, Universität Witten/Herdecke, Witten, Germany 5 Klinische Psychologie und Psychotherapie, Technische Universität Dresden, Dresden, Germany 2
The Behavioral Activation for Depression Scale (BADS) was developed to measure core concepts of behavioural activation for depression. A number of studies, mostly based on analogue samples, have provided initial support for the BADS. In the present study, we examined the psychometric properties of the German version of the scale more broadly, including change sensitivity and clinical treatment data. A mixed sample of students (N = 312) and depressed outpatients in partial remission undergoing cognitive-behavioural group treatment for depressive rumination (N = 59) was examined. To analyze construct validity, a set of theoretically relevant constructs such as perseverative thinking, distraction and mindfulness was also assessed. Results indicated good psychometric properties, additional evidence for construct validity of the total scale and subscales, and adequate fit of the data to the original factor structure. Furthermore, the BADS proved to be sensitive to changes in participants undergoing treatment for depression. Copyright © 2015 John Wiley & Sons, Ltd. Key Practitioner Message: • Behavioural activation (BA) is an effective treatment for patients suffering from unipolar depression. • The Behavioral Activation for Depression Scale (BADS) can be used to measure core elements of the BA treatment rationale. It is useful to track changes in activation within treatment. • The BADS is available in different languages and has shown to possess good psychometric properties. Keywords: Behavioural Activation, Depression, Scale Development, Measurement, Cognitive-behavioural Therapy
INTRODUCTION In the past decade, there has been renewed interest in the feasibility and efficacy of behavioural activation treatments for depression. Behavioural activation may be defined as a ‘structured, brief psychotherapeutic approach that aims to (a) increase engagement in adaptive activities (which often are those associated with the experience of pleasure and mastery), (b) decrease engagement in activities that maintain depression or increase risk for depression and (c) solve problems that limit access to reward or that maintain or increase aversive control’ (Dimidijan et al., 2011, p. 3).
*Correspondence to: Tobias Teismann, Clinical Psychology and Psychotherapy, Ruhr-Universität Bochum, Massenbergstraße 11, Bochum 44787, Germany. E-mail: [email protected]
Copyright © 2015 John Wiley & Sons, Ltd.
While therapists and researchers had seemed to prefer cognitive over behavioural interventions in depression in previous decades, Jacobson et al. (1996) showed in their seminal component analysis of cognitive-behavioural therapy for depression that there was no differential effectiveness of cognitive versus behavioural techniques—neither at post-treatment assessment nor at a two-year follow-up (Gortner, Gollan, Dobson, & Jacobsen, 1998). In response to this clear demonstration of the efficacy of behavioural activation, Martell, Addis, and Jacobson (2001) developed a full treatment model, called behavioral activation (BA), that employs activity scheduling to overcome deficits in positive reinforcement, and emphasizes the need to block avoidance behaviours such as rumination, social withdrawal and excessive reassurance seeking (cf. Ferster, 1973). Unlike early versions of BA (Lewinsohn, 1974), this manualized approach pays close attention to the unique environmental contingencies maintaining an
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218 individual’s depressed behaviour. Rather than focusing on the pleasantness of activities, it involves a detailed assessment of contingencies maintaining depressive behaviour, idiographic assessment of patients’ (behavioural) goals and the subsequent facilitation of goal-relevant behaviour—incorporating skills training, ruminationfocused strategies and contingency management. Dimidijan, Hollon, Dobson, et al. (2006) compared this version of BA with cognitive therapy, antidepressant medication and placebo in a large multisite randomized trial. In sum, all treatments performed similarly for mildly depressed patients; yet, BA and antidepressant medication outperformed cognitive therapy for moderate-to-severe depression, and BA evidenced a lower dropout rate than medication. Moreover, in cognitive therapy, a larger proportion of patients exhibited a pattern of extreme nonresponse by the end of the treatment than in BA (22% vs. 0%; Coffmann, Martell, Dimidjian, et al., 2007). Behavioural activation and its various versions have also gathered considerable empirical support in recent metaanalyses (Cuijpers, van Straten, & Warmerdam, 2007; Ekers, Richards, & Gilbody, 2008; Mazzucchelli, Kane, & Rees, 2009). Despite recent interest in behavioural treatments of depression, Manos, Kanter, and Bush (2010) identified a lack of good instruments to assess key variables of BA models. Frequently used measures, such as the Pleasant Events Schedule (MacPhillamy & Lewinsohn, 1982) or the Unpleasant Events Schedule (Lewinsohn & Talkington, 1979) not only confound the assessment of reinforcement with the assessment of mood, they also equate pleasantness with positive reinforcement and unpleasantness with punishment (Manos et al., 2010). Furthermore, aspects relevant to newer variants of BA, such as avoidance and rumination, are not captured by these instruments. Based on this background, Kanter et al. (2007) developed the Behavioral Activation for Depression Scale (BADS). The scale was generated through an exploratory factor analysis of a rationally derived set of items using an undergraduate students sample (N = 391) followed by a confirmatory factor analysis using a second undergraduate sample (N = 319; Kanter et al., 2007). The resulting scale consists of 25 items comprising four subscales: Activation (seven items: e.g., ‘I did something that was hard to do but it was worth it’ and ‘I engaged in a wide and diverse array of activities.’), Avoidance/Rumination (eight items: e.g., ‘I did things to avoid feeling sadness or other painful emotion’ and ‘Most of what I did was to escape from or avoid something unpleasant’), Work/School Impairment (five items: e.g., ‘There were certain things I needed to do that I didn’t do’ and ‘I was active, but did not accomplish any of my goals for the day’) and Social Impairment (five items: e.g., ‘I was not social, even though I had opportunities to be’ and ‘I did things to cut myself off from other people’). This factor structure was replicated in a community sample (N = 241) Copyright © 2015 John Wiley & Sons, Ltd.
with elevated depressive symptoms (Kanter, Rusch, Busch, & Sedivy, 2009), as well as in a Dutch community sample (N = 402; Raes et al., 2010). Adequate reliability and validity were reported for the BADS in all the abovementioned studies, with BADS scores showing predicted associations with criterion measures of depression, rumination, acceptance, avoidance and social support. Recently, a nine-item short form of the BADS—assessing activity and avoidance—has been introduced (BADS— Short Form; Manos, Kanter, & Luo, 2011). The current study aimed to examine the factor structure, reliability and construct validity, including change sensitivity and predictive accuracy, of the German version of the BADS within four independent non-clinical and clinical samples. The present study extends prior work on the BADS, as a sample of depressed patients diagnosed with a structured clinical interview was included for the first time. In order to explore the ability of the BADS to track changes over the course of treatment, data stemming from a randomized controlled trial comparing a cognitivebehavioural group treatment for depressive rumination with a wait-list control condition (Teismann, Hanning, von Brachel, & Willutzki, 2012; Teismann et al., 2014) were also analyzed. Finally, associations between BADS scores and further, theoretically interesting variables such as perseverative thinking, distraction and mindfulness were examined. Mindfulness is not actively trained in BA; however, in dealing with ruminative thoughts, Martell et al. (2001) recommend attending to actual sensory experiences in a mindful way. Mindfulness may be seen as the opposite of avoidance, rumination and disengagement and should therefore be positively associated with BA.
METHOD Participants The present study included participants from four independent samples recruited from several non-clinical settings and one clinical setting in Germany.
Student Samples (SSA) The student sample (SSA) comprised N = 312 participants from two samples assessed in Dresden (SSA-I: n = 85; SSA-II: n = 125) and in Bochum (SSA-III: n = 102). Of the total SSA, 67.0% (n = 209) were female. Age ranged from 17 to 68 years with a mean of 23.78 (standard deviation [SD] = 5.80). Thirty-two participants (15.2%) of the subsamples SSA-I and SSA-II were single and 178 (84.8%) lived in a romantic partnership. No respective information was available for sample SSA-III. The student subsamples did not differ regarding age and gender composition as well as depressed symptomatology and were Clin. Psychol. Psychother. 23, 217–225 (2016)
Behavioral Activation for Depression Scale therefore collapsed for all further analyses. Mean Beck Depression Inventory (BDI) score was M = 7.51 (SD = 6.46).
Depressed Outpatient Sample (DOS) The depressed outpatient sample (DOS) comprised N = 60 patients suffering from residual depression, who took part in a randomized controlled trial on the effectiveness of a rumination-focused cognitive-behavioural group treatment (Teismann et al., 2014). Since BADS data were missing from one person, sample size in the current analysis was reduced to n = 59. Seventy-two per cent (n = 43) of all patients were of female gender. Mean age was 47.12 years (SD = 11.77) ranging from 18 to 65. Forty patients (67.7%) were active, seven (11.9%) were students, six (10.2%) were retired and six (10.2%) were unemployed. Thirty-one patients (32.5%) were living in a romantic partnership, 13 (22%) were single, 14 (23.7%) were separated/divorced and 1 (1.7%) was widowed. Patients either suffered from major depressive disorder, recurrent, in partial remission (n = 46, 76.7%), or from major depressive disorder, single episode, in partial remission (n = 14, 23.3%). Thirty-eight per cent (n = 23) had suffered more than two episodes and 36.7% (n = 22) had suffered more than five episodes in their lifetime. Forty per cent (n = 24) was on stable antidepressant medication at study intake. The baseline BDI score was M = 20.10 (SD = 9.47). Prior to assessments, the participants were informed about the purpose of the study, the voluntary nature of their participation, data storage and security. They gave written and informed consent before participating. The study was approved by the Ethics Committee of the Faculty of Psychology at Ruhr-Universität Bochum.
Procedure The SSA was recruited in Dresden (SSA-I and SSA-II) and Bochum (SSA-III), Germany. Students were approached during lectures and courses in psychology. Furthermore, postings at various settings at the different universities promoted the study. Students received course credit for their participation. All questionnaires were presented in a paper–pencil format. The DOS was recruited in the Ruhr area in Germany. The depressed participants were recruited via letters to clinicians and articles about rumination and the treatment project in local newspapers and magazines. Research staff interviewed potential participants by telephone to screen for inclusion and exclusion criteria. Patients who passed the telephone screening were invited to a personal interview. During this session, an experienced, trained clinical psychologist administered the German version of the Structured Clinical Interview for Diagnostic and Statistical Manual (DSM)-IV diagnosis (First, Spitzer, Gibbon, & Williams, 1996). Patients were offered participation in the study, if Copyright © 2015 John Wiley & Sons, Ltd.
219 they met the following criteria: (a) DSM-IV (APA, 1994) criteria for major depressive disorder, recurrent, in partial remission (DSM-IV 296.35), and major depressive disorder, single episode, in partial remission (DSM-IV 296.25); (b) an initial score of ≥9 on the BDI-II (Beck, Steer, & Brown, 1996); (c) depression considered to be the most severe disorder if other comorbid disorders were present (i.e., severity status was the highest for the depressed symptomatology and participants reported to feel impaired mostly by their depression); (d) 18–65 years of age; (e) not meeting DSM-IV criteria for psychosis, mania, and current substance abuse and dependence; and (f) no other psychological treatment for the duration of the study. All questionnaires were presented in a paper–pencil form. Participants were randomly assigned to treatment (n = 31) or to a wait-list condition (n = 29). Patients on the wait-list received no treatment for at least 3 months. Since BADS data were missing from one person, sample size in the WAIT condition was reduced to n = 28. Cognitive-behavioral group treatment for depressive rumination(CBT-DR; Teismann et al., 2012) uses a combination of metacognitive therapy techniques, such as attention-training technique, disputation of positive and negative metacognitive beliefs, detached mindfulness (Wells, 2009) and strategies from behavioural activation, such as distraction, attending to sensory experiences, acceptance, problem-solving (Addis & Martell, 2004; Martell et al., 2001) to help patients overcoming depressive rumination. The treatment comprises 11 group sessions of 90 min. The treatment manual and the positive effects of the treatment on depression, rumination, metacognitive beliefs and depressive relapse have been described in detail elsewhere (Teismann et al., 2012, 2014). The main assessments reported in here were carried out at pretreatment/pre-wait-list (T1) and posttreatment/post-wait-list (T2).
Measures 1. BADS (Kanter et al., 2007; German version: Hoyer & Teismann, 2013). The BADS is a 25-item self-report measure with four subscales: Activation, Avoidance/Rumination, Work/School Impairment, and Social Impairment. Participants indicate to what extent each statement was true for them during the last week on a seven-point scale (not at all to completely). To score the BADS, items from all scales other than the Activation scale are reverse-coded, and then, all items are summed. To score the subscales, no items are reverse-coded. Adequate psychometric properties have been reported including good internal consistency, good test–retest reliability and evidence for construct validity for the total score and the subscale scores (Kanter et al., 2007, 2009). Clin. Psychol. Psychother. 23, 217–225 (2016)
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220 The German version of the BADS was developed by means of a translation–back-translation procedure according to relevant guidelines for the translation of psychometric instruments (Hambleton, 2001). First, the original English BADS was translated into German independently by the first and the last authors. Second, the two translated versions were translated back by two native speakers, and third, the two versions were evaluated, compared and approved—after minor wording adjustments. 2. BDI-II (Beck et al., 1996; German version: Hautzinger, Keller, & Kühner, 2006). The BDI is a 21-item selfreport measure that surveys depressive symptoms on a four-point severity scale. It has demonstrated good internal consistency (Cronbach’s α ≥ 0.89 across clinical and non-clinical samples) as well as adequate concurrent validity with clinical ratings of depression (Hautzinger et al., 2006). The internal consistency for the BDI-II was α = 0.87 in the SSA (SSA-I–SSA-III) and α = 0.89 in the clinical sample. 3. Beck Anxiety Inventory (BAI; Beck & Steer, 1993; German version: Margraf & Ehlers, 2007). The BAI is a 21-item questionnaire that measures cognitive and somatic symptoms of anxiety within the last week, with higher scores indicating increased anxiety (range = 0–63). Sample items include ‘unable to relax’ and ‘heart pounding or racing’. Good psychometric properties have been demonstrated among diverse samples (Beck & Steer, 1993). In the present study, internal consistency was α = 0.89 in the SSA (SSA-III) and α = 0.86 in the clinical sample. 4. Perseverative Thinking Questionnaire (PTQ; Ehring et al., 2011). The PTQ is a 15-item self-report measure that assesses characteristics of ruminative thinking (e. g., repetitive, difficult to disengage from, unproductive, capturing mental capacity). All items are answered on a five-point scale ranging from 0 (never) to 4 (almost always). The PTQ total score has an excellent internal consistency: Cronbach’s α = 0.95 (Ehring et al., 2011). Accordingly, internal consistency was good in the SSA (SSA-III) α = 0.95 and in the clinical sample α = 0.91. 5. Response Styles Questionnaire (RSQ; NolenHoeksema & Morrow, 1991; German version: Kühner, Huffziger, & Nolen-Hoeksema, 2007). The RSQ is a 32item instrument that was designed to measure the way in which an individual typically responds to feelings of depression or sad mood. Participants are asked to rate 21 ruminative responses (e.g., ‘I go away by myself and think about why I feel this way’) and 11 distracting responses (e.g., ‘I go to a favorite place to get my mind off my feelings’) to depressed mood on a four-point scale ranging from 1 (almost never) to 4 (almost always). The two subscales have been shown to have acceptable internal consistency (Kühner et al., Copyright © 2015 John Wiley & Sons, Ltd.
2007). Accordingly, internal consistency of the rumination subscale (SSA, SSA-III, α = 0.79; clinical sample, α = 88) and the distraction subscale (SSA, SSA-III, α = 0.84; clinical sample, α = 0.81) was good in the present sample. 6. Mindfulness Attention Awareness Scale (MAAS; Brown & Ryan, 2003; German version: Michalak, Heidenreich, Ströhle, & Nachtigall, 2008). The MAAS is a 15-item self-report instrument measuring the general tendency to be attentive to and aware of presentmoment experiences in daily life (e.g., ‘I find it difficult to stay focused on what’s happening in the present’; ‘I could be experiencing some emotion and not be conscious of it until sometime later’). It consists of a single factor and thus yields a single total score. In the present study, the MAAS has an internal consistency of α = 0.90 in the SSA (SSA-III) and of α = 0.89 in the clinical sample.
STATISTICAL ANALYSES SPSS 20 for Windows (IBM, Armonk, NY, USA) was used for the statistical analyses as well as AMOS 20 (IBM, Armonk, NY, USA), which was used to carry out confirmatory factor analysis. Internal consistencies of the derived scales were determined by calculating Cronbach’s α. Construct validity was evaluated for samples 1 and 2 by calculating Pearson correlation coefficients between the scales of the BADS and the other scales used in this study. Change sensitivity was assessed by analyzing differences between the treatment and the wait-list group at the post-treatment-wait assessment using one-way analyses of covariance, with pretreatment scores as covariate. Furthermore, significant pretreatment to post-treatment changes within the groups were analyzed by repeatedmeasure analyses of variances (ANOVAs). Within-group effect sizes (pre to post) were calculated according to the recommendation of Cohen (1988). All of the latter analyses were intent-to-treat analysis, with last data carried forward. Post data of two patients (6.5%) in the treatment group and four patients (14.3%) in the wait-list condition had to be inserted (cf. Teismann et al., 2014). Test–retest reliability, assessed with Pearson correlation coefficients, was established in participants of the wait-list condition, who returned for the post-wait assessment 3 to 5 months after the pre-wait assessment.
RESULTS Confirmatory Factor Analysis The hypothesized four-factor model of the German version of the BADS was investigated by confirmatory factor Clin. Psychol. Psychother. 23, 217–225 (2016)
Behavioral Activation for Depression Scale analysis using maximum-likelihood estimation and performing Bollen–Stine bootstrap because of nonnormality of the data. The sample consisted of n = 371 participants (n = 59 of them patients and n = 312 students). The overly sensitive chi-square (Tabachnick & Fidell, 2001) reached significance (χ 2 = 1031.875; df = 269; p < 0.001) indicating a poor model fit. Additional fit indices were regarded following the recommendations given by Hu and Bentler (1999). The root mean square error of approximation (RMSEA; which should be below 0.08) of 0.088, 90% confidence interval [0.082, 0.093] with p < 0.01, does not support an acceptable fit. The comparative fit index (CFI; which should be 0.95 or above) of 0.779 does not support model fit either. Only the standardized root mean residual (SRMR; a value of 0.11 or below indicating acceptable fit) of 0.0856 supports model fit. Therefore, modification indices were inspected. Following the highest modification index (MI) of MI = 128 864, fit indices were calculated taking into account the suggested
221 covariance between variables e11 and e12. Again, chisquare reached significance (χ 2 = 879 652; df = 268; p < 0.001). The RMSEA of 0.079 indicates moderate model fit as does the SRMR of 0.081. Still, the CFI of 0.822 does not indicate an adequate model fit. Figure 1 presents the completely standardized factor solution of the confirmatory factor analysis. The residuals (expressed as covariances) of the indicators were mostly in the small-to-medium range (0.00–0.78), suggesting that the model explained most indicators well. Most factor loadings are in an acceptable range (λ = 0.34–78) except for items 23 (λ = 0.16) and item 25 (λ = 0.17). The latent variable Activation is associated negatively with Avoidance/Rumination expressed by a covariance of 0.46, Work/School Impairment expressed by a covariance of 0.72 and Social Impairment expressed by a covariance of 0.31, while the latter three latent variables were associated positively (expressed by covariances ranging from 0.43 to 0.63).
Figure 1. Completely standardized factor solution of the confirmatory factor analysis. Act = Activation; Av_Ru = Avoidance/Rumination; Work_School_Imp = Work/School Impairment; Soc_Imp = Social Impairment; BADS = items of the Behavioral Activation for Depression scale
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Scale Properties Internal consistency was assessed using Cronbach’s α. The 25-item BADS had an overall α of 0.85 in both samples, respectively. The internal consistencies of the subscales were Activation (seven items, SSA: α = 0.76; clinical sample: α = 0.79); Avoidance/Rumination (eight items, SSA: α = 0.76; clinical sample: α = 0.76); Work/School Impairment (five items, SSA: α = 0.63; clinical sample: α = 0.60); and Social Impairment (five items, SSA: α = 0.81; clinical sample: α = 0.83). Having taken item number into account, all Cronbach’s α were sufficient (cf. Cortina, 1993). Test–retest reliability, assessed with Pearson correlation coefficients, was established on the 28 depressed participants of the wait-list condition (DOS), who returned for the post-wait assessment (T2) 3 to 5 months (M = 5.04, SD = 1.5 months) after the pre-wait assessment (T1). The BADS total scale (r = 0.51, p < 0.01), as well as the BADS subscales demonstrated sufficient test–retest reliability: Activation (r = 0.67, p < 0.001); Avoidance/Rumination (r = 0.62, p < 0.001); Work/School Impairment (r = 0.66, p < 0.001); and Social Impairment (r = 0.57, p < 0.01).
both samples. Only BADS Work/School Impairment was unrelated to depression severity in the DOS. Anxiety scores (BAI) were correlated with BADS scores in the SSA, but unrelated in the DOS. Regarding perseverative thinking scores (PTQ), correlations were in the expected direction in the SSA. In the DOS, perseverative thinking was strongly associated with the BADS Avoidance/ Rumination scale and moderately correlated with increased BADS Social Impairment scores. Depressive rumination (RSQ—Rumination Scale) was associated with BADS Avoidance/Rumination scores in both samples and with BADS Social Impairment in the DOS. Habitual distraction in response to depressed mood (RSQ— Distraction Scale) was associated with increased BADS Activation scores in both samples. Furthermore, increased distraction was associated with reduced Work/School Impairment in the DOS. Increased mindfulness (MAAS) was associated with reduced Avoidance/Rumination, Work/School Impairment and Social Impairment in the SSA, but was unrelated to BADS scores in the DOS.
Change Sensitivity Construct validity Table 1 presents the correlations between the BADS and its subscales on the one hand, and the criterion measures on the other hand. Regarding severity of depressive symptoms (BDI-II), correlations with the BADS total score and most subscales were in the expected direction—in Table 1.
In order to explore the ability of the BADS to track changes over the course of treatment, data steaming from the randomized controlled trial comparing CBT-DR with a wait-list control condition (Teismann et al., 2014) were analyzed. Table 2 shows participants’ scores at pretreatment wait and post-treatment wait as well as treatment effects. At pretreatment, ANOVA indicated that there were no
Correlations between BADS total/subscales and additional measures BDI
BAI
PTQ
RSQ-R
RSQ-D
MAAS
BADS total Student sample Outpatient sample
0.539** 0.429**
0.434** 0.220
0.466** 0.367**
0.231* 0.360**
0.214* 0.311*
0.376** 0.174
Activation Student sample Outpatient sample
0.246** 0.256*
0.261** 0.023
0.251* 0.068
0.132 0.161
0.307** 0.384**
0.110 0.064
Avoidance/Rumination Student sample Outpatient sample
0.512** 0.374**
0.472** 0.335*
0.472** 0.531**
0.216* 0.342**
0.057 0.080
0.424** 0.235
Work/School Impairment Student sample Outpatient sample
0.353** 0.182
0.236* 0.017
0.239* 0.104
0.190 0.207
0.151 0.316*
0.199* 0.108
Social Impairment Student sample Outpatient sample
0.429** 0.382**
0.183 0.201
0.312** 0.286*
0.102 0.303*
0.098 0.139
0.298** 0.204
BADS = Behavioral Activation for Depression Scale, BAI = Beck Anxiety Inventory; BDI = Beck Depression Inventory; MAAS = Mindfulness Attention and Awareness Scale; PTQ = Perseverative Thinking Scale; RSQ-D = Response Styles Questionnaire—Distraction Scale; RSQ-R = Response Styles Questionnaire—Rumination Scale. **p < 0.01, *p < 0.05
Copyright © 2015 John Wiley & Sons, Ltd.
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Table 2. BADS outcome measures at each assessment CBT-DR (n = 31)
WAIT (n = 28)
Measure/time
M
SD
M
SD
Group effect†
BADS—total Pretreatment Post-treatment
87.45 96.61
21.81 18.84
83.71 84.78
16.79 20.37
F(1, 57) = 0.54 F(1, 56) = 5.69*
BADS—Activation Pretreatment Post-treatment
21.42 24.29
8.28 8.60
22.96 23.11
6.88 6.69
F(1, 57) = 0.59 F(1, 56) = 2.24
BADS—Avoidance/Rumination Pretreatment 22.16 Post-treatment 17.77
9.13 9.66
23.33 23.14
7.71 9.21
F(1, 57) = 0.27 F(1, 56) = 5.91*
BADS—Work/School Impairment Pretreatment 13.26 Post-treatment 11.09
5.88 7.19
10.96 11.28
5.95 6.09
F(1, 57) = 2.21 F(1, 56) = 2.84
BADS—Social Impairment Pretreatment Post-treatment
8.29 7.18
10.18 10.41
6.25 6.99
F(1, 57) = 0.09 F(1, 56) = 4.77*
10.77 7.29
†
At pretreatment, the group effect is based on a one-way analysis of variance. At post-treatment, the group effect is based on a one-way analysis of covariance, with pretreatment scores as the covariate. CBT-DR = cognitive-behavioral group treatment for depressive rumination; BADS = Behavioral Activation for Depression scale; M = mean; SD = standard deviation. *p < 0.05
significant differences between the groups on any BADS. At post-treatment, a series of one-way analyses of covariances indicated that CBT-DR was superior to WAIT regarding the BADS total score as well as the subscales Avoidance/ Rumination and Social Impairment. No group differences were observable on the BADS subscales Activity and School/Work Impairment. Repeated-measure ANOVAs were used to assess pretreatment-wait to post-treatment-wait change. CBTDR was associated with statistically significant changes of the BADS total [F(1, 30) = 13.11, p < 0.001; d = 0.47] as well as on all BADS subscales [BADS-Activity, F(1, 30) = 5.88, p < 0.05, d = 0.38; BADS-Avoidance/Rumination, F (1, 30) = 12.28, p < 0.001, d = 0.52; BADS-Work Impairment, F(1, 30) = 7.36, p < 0.01, d = 0.36; BADS-Social Impairment, F(1, 30) = 7.04, p < 0.01, d = 0.47]. No improvements at all were observed in the WAIT condition.
DISCUSSION In the present study, the factor structure, reliability, construct validity and change sensitivity of the German version of the BADS were investigated. Using confirmatory factor analysis, the four-factor structure reported by Kanter et al. (2007, 2009) for the original English version of the BADS provided a sufficient fit to the German version of the BADS. The original model was slightly adjusted following the highest modification index. Allowing for covariation of error variables e11 and e12 Copyright © 2015 John Wiley & Sons, Ltd.
resulted in a moderate-to-good model fit for the two fit indices (RMSEA and SRMR), while the CFI score was less impressive. With regard to contents, items 11 and 12 show a similarity that distinguishes them from the other indicators of the latent variable Activation. While all other five items refer directly to activation itself, items 11 (‘I did things even though they were hard because they fit in with long-term goals for myself’) and 12 (‘I did something that was hard to do but it was worth it’) additionally contain the denotation of bringing oneself to something even though it might be difficult (‘hard’). A model fit similar to our results has been described by Kanter et al. (2009) in their study on a community sample with elevated depression scores. Following their argumentation, it might as well be true for our results that the moderate sample size might have contributed to a fit less than perfect. In addition, Kanter et al. (2009) suggest considering investigating alternate model structures. The point that using a combined sample of a majority of students and a minority of patients for the confirmatory factor analysis might be responsible for a fit less than perfect can be ruled out; a tentative analysis only using the SSA provided equal results. Two items (items 23 and 25) of the BADS had quite low factor loadings. To ensure comparability with international studies, they were nonetheless retained in the questionnaire—a tentative evaluation omitting those items did not lead to any change in model fit. Internal consistencies and test–retest reliability were acceptable. (Only for one subscale the low number of items is presumably at the expense of comparably lower Clin. Psychol. Psychother. 23, 217–225 (2016)
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224 internal consistency.) This is especially interesting, since the test–retest interval in the current study was longer than in any study before and since BA is generally viewed as a state rather than a trait characteristic. Construct validity of the BADS was supported by expected associations between BADS total score and subscales and a set of relevant criterion measures of depression, perseverative thinking, depressive rumination, distraction and mindfulness. As in previous studies, depressive symptomatology was consistently negatively associated with BADS scores (Kanter et al., 2007, 2009; Raes et al., 2010). Associations with anxiety were less consistent (cf. Raes et al., 2010). As expected, it was found that the BADS-Avoidance/Rumination subscale correlated significantly with perseverative thinking and depressive rumination, respectively. Interestingly, there were also substantial associations between the BADS-Social Impairment subscale and perseverative thinking. This fits with earlier work showing that habitual rumination has a marked adverse effect on social relationships (e.g., Lam et al., 2003; Müller et al., 2012). Furthermore, it was shown that distractive responses to depressed mood, as assessed with the RSQ, are significantly correlated with increased activation, as assessed with the BADS-Activation subscale. Making use of distraction, in response to ruminative thinking as well as depressed mood, is actively trained in behavioural activation (Martell et al., 2001) and has repeatedly been shown to be an effective strategy for the short-termed regulation of emotional distress (Nolen-Hoeksema, Wisco, & Lyubomirsky, 2008). Taken together, most associations were stronger in the SSA than in the DOS. Therefore, future studies should reassess the clinical use of the BADS in larger samples of depressed patients. Different from what was expected, associations between mindfulness and BADS scores were only detected in the SSA but not in the outpatient sample. One possible explanation for this differential finding is that non-depressed students who engage in goal-relevant activities manage to do so in a mindful way, i.e., attentionally focused on what they are doing. Some (but not all) depressed outpatients, in contrast, may be more susceptible to internal and external distractions, and even though they become more active, they may have difficulties in focusing their attention on the task at hand. Finally, the BADS proved to be useful in tracking—even small to moderate—changes in activation reported by patients undergoing treatment. In general, CBT-DRs’ effect on facets of BA, such as an increase in Activity, and decreases in Avoidance behaviour (including Rumination) and Work and Social Impairment, go well with the therapeutic focus of the treatment. Although strategies like structured activity monitoring and scheduling were not an integral part of the treatment, patients were frequently reminded to become active when (and instead of) ruminating. Furthermore, strategies in dealing with rumination, such as distraction, attention to sensory experiences and Copyright © 2015 John Wiley & Sons, Ltd.
problem-solving (Martell et al., 2001) constituted a central part of the treatment. It is, however, unclear to what extent reductions in social impairments are due to these specific inputs of the therapy session. They may also have been instigated by the group setting itself: Thus, joining the group session itself may already significantly counteract social withdrawal behaviour. Furthermore, experiencing oneself in an interpersonal context may have helped to overcome feelings of social insufficiency and isolation. A set of further intriguing research questions would become testable by the use of the BADS in future studies on psychotherapy of depression: to what extent would different treatments for depression differ from each other in their effect on BA? do changes in activation predate and predict changes in depression? and finally, what is the importance of BA, as assessed with the BADS, for the prediction of depressive relapse and recurrences? Several limitations have to be considered, when interpreting the results. First, the sample size of the depressed outpatients was rather small. Future studies should attempt to utilize larger patient samples making it possible to evaluate the factor structure of the BADS separately for different clinical and non-clinical samples. Second, an acceptable model fit was only established after allowing for a—theoretically underpinned—covariation between two error variables. Therefore, a cross-validation of the here-described factor structure is necessary. In sum, the present results are largely consistent with those of previous research with the original English version of the BADS (Kanter et al., 2007, 2009). They extend previous research in several important ways by broadening information on reliability and construct validity (including theoretically relevant associations of the BADS, which had previously not been investigated) and employing a sample of depressed patients undergoing treatment.
ACKNOWLEDGEMENTS The study was funded by a grant from the German Research Society (Deutsche Forschungsgemeinschaft, DFG) to UW (WI 1106/12-1).
REFERENCES Addis, M. E., & Martell, C. R. (2004). Overcoming depression one step at a time. New York: New Harbinger Press. Beck, A.T., & Steer, R.A. (1993). Beck Anxiety Inventory Manual. San Antonio: Harcourt Brace and Company. Beck, A.T., Steer R.A., & Brown, G.K. (1996). Beck Depression Inventory – Second Edition. San Antonio: The Psychological Cooperation. Brown, K.W., & Ryan, R.M. (2003). The benefits of being present: mindfulness and its role in psychological well-being. Journal of Personality and Social Psychology, 84, 822–848.
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Behavioral Activation for Depression Scale Coffmann, S.J., Martell, C.R., Dimidjian S., Gallop R., & Hollon S.D. (2007). Extreme nonresponse in cognitive therapy: Can behavioural activation succeed where cognitive therapy fails? Journal of Consulting and Clinical Psychology, 75, 531–541. Cohen, J. (1988). Statistical power analysis for the behavioural sciences. Hillsdale: Erlbaum. Cortina, J. M. (1993). What is coefficient alpha? An examination of theory and applications. Journal of Applied Psychology, 78, 98–104. Cuijpers, P., van Straten, A., & Warmerdam, L. (2007). Behavioural activation treatments of depression: A meta-analysis. Clinical Psychology Review, 27, 318–326. Dimidijan, S., Barrera, M., Martell, C., Munoz, R.F., & Lewinsohn, P.M. (2011). The origins and current status of behavioral activation treatments for depression. Annual Review of Clinical Psychology, 7, 1–38. Dimidijan, S., Hollon, S.D., Dobson, K.S., Schmaling, K.B., Kohlenberg, R.J., Addis, M.E., … Jacobson, N.S. (2006). Randomized trial of behavioral activation, cognitive therapy, and antidepressed mediation in the acute treatment of adults with major depression. Journal of Consulting and Clinical Psychology, 74, 658–670. Ehring, T., Zetsche, U., Weidacker, K., Wahl, K., Schönfeld, S., & Ehlers, A. (2011). The Perseverative Thinking Questionnaire (PTQ): Validation of a content-independent measure of repetitive negative thinking. Journal of Behavior Therapy and Experimental Psychiatry, 42, 225–232. Ekers, D., Richards, D., & Gilbody, S. (2008). A meta-analysis of randomized trials of behavioural treatment of depression. Psychological Medicine, 38, 611–623. Ferster, C. (1973). A functional analysis of depression. American Psychologist, 28, 857–870. First, M.B., Spitzer, R.L., Gibbon, M., & Williams, J.B.W. (1996). Structured Clinical Interview for DSM-IVAxis I Disorders. Washington, D.C.: American Psychiatric Press. Gortner, E.T., Gollan, J.K., Dobson, K.S., & Jacobsen, N.S. (1998). Cognitive-behavioral treatment for depression: Relapse prevention. Journal of Consulting and Clinical Psychology, 66, 377–384. Hambleton, R. K. (2001). The next generation of the ITC test translation and adaptation guidelines. European Journal of Psychological Assessment, 17, 164–172. Hautzinger, M., Keller, F., & Kühner, C. (2006). BDI-II. Frankfurt a.M.: Harcourt Test Service. Hoyer, J., & Teismann, T. (2013). Behavioral activation for depression scale – German version. Unpublished manuscript, Institute for Clinical psychology and Psychotherapy, Germany: Technische Universität Dresden. Hu, L., & Bentler, P. M. (1999). Cutoff criteria for fit indexes in covariance structure analysis: Conventional criteria versus new alternatives. Structural Equation Modeling: A Multidisciplinary Journal, 6, 1–55. Jacobson, N.S., Dobson, K.S., Truax, P.A., Addis, M.E., Koerner, K., Gollan, J.K., … Prince, S.E. (1996). A component analysis of cognitive-behavioral treatment for depression. Journal of Consulting and Clinical Psychology, 64, 295–304. Kanter, J.W., Mulick, P.S., Busch, A.W., Berlin, K.S., & Martell, C. R. (2007). The behavioral activation for depression scale (BADS): Psychometric properties and factor structure. Journal of Psychopathology and Behavioral Assessment, 29, 191–202. Kanter, J.W., Rusch, L.C., Busch, A.M., & Sedivy, S.K. (2009). Validation of the behavioral activation for depression scale (BADS) in a community sample with elevated depressive symptoms. Journal of Psychopathology and Behavioral Assessment, 31, 36–42.
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225 Kühner, C., Huffziger, S., & Nolen-Hoeksema, S. (2007). RSQ-D. Response Styles Questionnaire – Deutsche Version. Göttingen: Hogrefe. Lam, D., Schuck, N., Smith, N., Farmer, A., & Checkley, S. (2003). Response style, interpersonal difficulties and social functioning in major depressive disorder. Journal of Affective Disorders, 75, 279–283. Lewinsohn, P. (1974). A behavioral approach to depression. In R. J. Friedman & M.M. Katz (Eds.), Psychology of depression: Contemporary theory and research. Oxford: Wiley & Sons. Lewinsohn, P., & Talkington J. (1979). Studies on the measurement of unpleasant events and relations with depression. Applied Psychological Measurement, 3, 83–101. MacPhillamy, D.J., & Lewinsohn, P. (1982). The pleasant events scale: Studies on reliability, validity and scale intercorrelations. Journal of Consulting and Clinical Psychology, 50, 362–380. Manos, R.C., Kanter, J.W., & Bush, A.M. (2010). A critical review of assessment strategies to measure the behavioral activation model of depression. Clinical Psychology Review, 30, 547–561. Manos, R.C., Kanter, J.W., & Luo, W. (2011). The behavioral activation scale for depression – short form: Development and validation. Behavior Therapy, 42, 726–739. Margraf, J., & Ehlers, A. (2007). BAI: Beck Angst-Inventar. Manual. Deutsche Bearbeitung. Frankfurt am Main: Harcourt Test Services GmbH. Martell, C. R., Addis, M. E., & Jacobson, N. S. (2001). Depression in context. Strategies for guided action. New York: W.W.Norton. Mazzucchelli, T., Kane, R., & Rees, C. (2009). Behavioral activation treatments for depression in adults. Clinical Psychology: Science and Practice, 16, 383–411. Michalak, J., Heidenreich, T., Ströhle, G., & Nachtigall, C. (2008). Die deutsche Version der Mindful Attention and Awareness Scale (MAAS): Psychometrische Befunde zu einem Achtsamkeitsfragebogen. Zeitschrift für Klinische Psychologie und Psychotherapie, 37, 200–208. Müller, D., Teismann, T., Havemann, B., Michalak, J. & Seehagen, S. (2012). Ruminative thinking as a predictor of postpartum mother-infant bonding. Cognitive Therapy and Research, 37, 89-96. Nolen-Hoeksema, S., & Morrow, J. (1991). A prospective study of depression and posttraumatic stress symptoms after a natural disaster: The 1989 Loma Prieta earthquake. Journal of Personality and Social Psychology, 61, 115–121. Nolen-Hoeksema, S., Wisco, B. E., & Lyubomirsky, S. (2008). Rethinking rumination. Perspectives on Psychological Science, 5, 400–424. Raes, F., Hoes, D., van Gucht, D., Kanter, J.W., & Hermans, D. (2010). The Dutch version of the behavioral activation for depression scale (BADS): Psychometric properties and factor structure. Journal of Behavior Therapy and Experimental Psychiatry, 41, 246–250. Tabachnick & Fidell (2001). Using multivariate statistics. Boston: Pearson. Teismann, T., Hanning, S., von Brachel, R., & Willutzki, U. (2012). Kognitive Verhaltenstherapie depressiven Grübelns. Berlin: Springer. Teismann, T., von Brachel, R., Hanning, S., Grillenberger, M., Hebermehl, L., Hornstein, I., … Willutzki, U. (2014). A randomized controlled trial on the effectiveness of a ruminationfocused group treatment for residual depression. Psychotherapy Research 24, 80–90. Wells, A. (2009). Metacognitive therapy for anxiety and depression. New York: Guilford Press.
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The German Version of the Behavioral Activation for Depression Scale (BADS): A Psychometric and Clinical Investigation Tobias Teismann,1* Andrea Ertle,2 Nadine Furka,3 Ulrike Willutzki4 and Juergen Hoyer5 1
Department of Clinical Psychology and Psychotherapy, Ruhr-Universität Bochum, Bochum, Germany Institute of Psychology, Humboldt-Universität zu Berlin, Berlin, Germany 3 Institute of Clinical Psychology and Psychotherapy, Technische Universität Dresden, Dresden, Germany 4 Department for Psychology and Psychotherapy, Universität Witten/Herdecke, Witten, Germany 5 Klinische Psychologie und Psychotherapie, Technische Universität Dresden, Dresden, Germany 2
The Behavioral Activation for Depression Scale (BADS) was developed to measure core concepts of behavioural activation for depression. A number of studies, mostly based on analogue samples, have provided initial support for the BADS. In the present study, we examined the psychometric properties of the German version of the scale more broadly, including change sensitivity and clinical treatment data. A mixed sample of students (N = 312) and depressed outpatients in partial remission undergoing cognitive-behavioural group treatment for depressive rumination (N = 59) was examined. To analyze construct validity, a set of theoretically relevant constructs such as perseverative thinking, distraction and mindfulness was also assessed. Results indicated good psychometric properties, additional evidence for construct validity of the total scale and subscales, and adequate fit of the data to the original factor structure. Furthermore, the BADS proved to be sensitive to changes in participants undergoing treatment for depression. Copyright © 2015 John Wiley & Sons, Ltd. Key Practitioner Message: • Behavioural activation (BA) is an effective treatment for patients suffering from unipolar depression. • The Behavioral Activation for Depression Scale (BADS) can be used to measure core elements of the BA treatment rationale. It is useful to track changes in activation within treatment. • The BADS is available in different languages and has shown to possess good psychometric properties. Keywords: Behavioural Activation, Depression, Scale Development, Measurement, Cognitive-behavioural Therapy
INTRODUCTION In the past decade, there has been renewed interest in the feasibility and efficacy of behavioural activation treatments for depression. Behavioural activation may be defined as a ‘structured, brief psychotherapeutic approach that aims to (a) increase engagement in adaptive activities (which often are those associated with the experience of pleasure and mastery), (b) decrease engagement in activities that maintain depression or increase risk for depression and (c) solve problems that limit access to reward or that maintain or increase aversive control’ (Dimidijan et al., 2011, p. 3).
*Correspondence to: Tobias Teismann, Clinical Psychology and Psychotherapy, Ruhr-Universität Bochum, Massenbergstraße 11, Bochum 44787, Germany. E-mail: [email protected]
Copyright © 2015 John Wiley & Sons, Ltd.
While therapists and researchers had seemed to prefer cognitive over behavioural interventions in depression in previous decades, Jacobson et al. (1996) showed in their seminal component analysis of cognitive-behavioural therapy for depression that there was no differential effectiveness of cognitive versus behavioural techniques—neither at post-treatment assessment nor at a two-year follow-up (Gortner, Gollan, Dobson, & Jacobsen, 1998). In response to this clear demonstration of the efficacy of behavioural activation, Martell, Addis, and Jacobson (2001) developed a full treatment model, called behavioral activation (BA), that employs activity scheduling to overcome deficits in positive reinforcement, and emphasizes the need to block avoidance behaviours such as rumination, social withdrawal and excessive reassurance seeking (cf. Ferster, 1973). Unlike early versions of BA (Lewinsohn, 1974), this manualized approach pays close attention to the unique environmental contingencies maintaining an
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218 individual’s depressed behaviour. Rather than focusing on the pleasantness of activities, it involves a detailed assessment of contingencies maintaining depressive behaviour, idiographic assessment of patients’ (behavioural) goals and the subsequent facilitation of goal-relevant behaviour—incorporating skills training, ruminationfocused strategies and contingency management. Dimidijan, Hollon, Dobson, et al. (2006) compared this version of BA with cognitive therapy, antidepressant medication and placebo in a large multisite randomized trial. In sum, all treatments performed similarly for mildly depressed patients; yet, BA and antidepressant medication outperformed cognitive therapy for moderate-to-severe depression, and BA evidenced a lower dropout rate than medication. Moreover, in cognitive therapy, a larger proportion of patients exhibited a pattern of extreme nonresponse by the end of the treatment than in BA (22% vs. 0%; Coffmann, Martell, Dimidjian, et al., 2007). Behavioural activation and its various versions have also gathered considerable empirical support in recent metaanalyses (Cuijpers, van Straten, & Warmerdam, 2007; Ekers, Richards, & Gilbody, 2008; Mazzucchelli, Kane, & Rees, 2009). Despite recent interest in behavioural treatments of depression, Manos, Kanter, and Bush (2010) identified a lack of good instruments to assess key variables of BA models. Frequently used measures, such as the Pleasant Events Schedule (MacPhillamy & Lewinsohn, 1982) or the Unpleasant Events Schedule (Lewinsohn & Talkington, 1979) not only confound the assessment of reinforcement with the assessment of mood, they also equate pleasantness with positive reinforcement and unpleasantness with punishment (Manos et al., 2010). Furthermore, aspects relevant to newer variants of BA, such as avoidance and rumination, are not captured by these instruments. Based on this background, Kanter et al. (2007) developed the Behavioral Activation for Depression Scale (BADS). The scale was generated through an exploratory factor analysis of a rationally derived set of items using an undergraduate students sample (N = 391) followed by a confirmatory factor analysis using a second undergraduate sample (N = 319; Kanter et al., 2007). The resulting scale consists of 25 items comprising four subscales: Activation (seven items: e.g., ‘I did something that was hard to do but it was worth it’ and ‘I engaged in a wide and diverse array of activities.’), Avoidance/Rumination (eight items: e.g., ‘I did things to avoid feeling sadness or other painful emotion’ and ‘Most of what I did was to escape from or avoid something unpleasant’), Work/School Impairment (five items: e.g., ‘There were certain things I needed to do that I didn’t do’ and ‘I was active, but did not accomplish any of my goals for the day’) and Social Impairment (five items: e.g., ‘I was not social, even though I had opportunities to be’ and ‘I did things to cut myself off from other people’). This factor structure was replicated in a community sample (N = 241) Copyright © 2015 John Wiley & Sons, Ltd.
with elevated depressive symptoms (Kanter, Rusch, Busch, & Sedivy, 2009), as well as in a Dutch community sample (N = 402; Raes et al., 2010). Adequate reliability and validity were reported for the BADS in all the abovementioned studies, with BADS scores showing predicted associations with criterion measures of depression, rumination, acceptance, avoidance and social support. Recently, a nine-item short form of the BADS—assessing activity and avoidance—has been introduced (BADS— Short Form; Manos, Kanter, & Luo, 2011). The current study aimed to examine the factor structure, reliability and construct validity, including change sensitivity and predictive accuracy, of the German version of the BADS within four independent non-clinical and clinical samples. The present study extends prior work on the BADS, as a sample of depressed patients diagnosed with a structured clinical interview was included for the first time. In order to explore the ability of the BADS to track changes over the course of treatment, data stemming from a randomized controlled trial comparing a cognitivebehavioural group treatment for depressive rumination with a wait-list control condition (Teismann, Hanning, von Brachel, & Willutzki, 2012; Teismann et al., 2014) were also analyzed. Finally, associations between BADS scores and further, theoretically interesting variables such as perseverative thinking, distraction and mindfulness were examined. Mindfulness is not actively trained in BA; however, in dealing with ruminative thoughts, Martell et al. (2001) recommend attending to actual sensory experiences in a mindful way. Mindfulness may be seen as the opposite of avoidance, rumination and disengagement and should therefore be positively associated with BA.
METHOD Participants The present study included participants from four independent samples recruited from several non-clinical settings and one clinical setting in Germany.
Student Samples (SSA) The student sample (SSA) comprised N = 312 participants from two samples assessed in Dresden (SSA-I: n = 85; SSA-II: n = 125) and in Bochum (SSA-III: n = 102). Of the total SSA, 67.0% (n = 209) were female. Age ranged from 17 to 68 years with a mean of 23.78 (standard deviation [SD] = 5.80). Thirty-two participants (15.2%) of the subsamples SSA-I and SSA-II were single and 178 (84.8%) lived in a romantic partnership. No respective information was available for sample SSA-III. The student subsamples did not differ regarding age and gender composition as well as depressed symptomatology and were Clin. Psychol. Psychother. 23, 217–225 (2016)
Behavioral Activation for Depression Scale therefore collapsed for all further analyses. Mean Beck Depression Inventory (BDI) score was M = 7.51 (SD = 6.46).
Depressed Outpatient Sample (DOS) The depressed outpatient sample (DOS) comprised N = 60 patients suffering from residual depression, who took part in a randomized controlled trial on the effectiveness of a rumination-focused cognitive-behavioural group treatment (Teismann et al., 2014). Since BADS data were missing from one person, sample size in the current analysis was reduced to n = 59. Seventy-two per cent (n = 43) of all patients were of female gender. Mean age was 47.12 years (SD = 11.77) ranging from 18 to 65. Forty patients (67.7%) were active, seven (11.9%) were students, six (10.2%) were retired and six (10.2%) were unemployed. Thirty-one patients (32.5%) were living in a romantic partnership, 13 (22%) were single, 14 (23.7%) were separated/divorced and 1 (1.7%) was widowed. Patients either suffered from major depressive disorder, recurrent, in partial remission (n = 46, 76.7%), or from major depressive disorder, single episode, in partial remission (n = 14, 23.3%). Thirty-eight per cent (n = 23) had suffered more than two episodes and 36.7% (n = 22) had suffered more than five episodes in their lifetime. Forty per cent (n = 24) was on stable antidepressant medication at study intake. The baseline BDI score was M = 20.10 (SD = 9.47). Prior to assessments, the participants were informed about the purpose of the study, the voluntary nature of their participation, data storage and security. They gave written and informed consent before participating. The study was approved by the Ethics Committee of the Faculty of Psychology at Ruhr-Universität Bochum.
Procedure The SSA was recruited in Dresden (SSA-I and SSA-II) and Bochum (SSA-III), Germany. Students were approached during lectures and courses in psychology. Furthermore, postings at various settings at the different universities promoted the study. Students received course credit for their participation. All questionnaires were presented in a paper–pencil format. The DOS was recruited in the Ruhr area in Germany. The depressed participants were recruited via letters to clinicians and articles about rumination and the treatment project in local newspapers and magazines. Research staff interviewed potential participants by telephone to screen for inclusion and exclusion criteria. Patients who passed the telephone screening were invited to a personal interview. During this session, an experienced, trained clinical psychologist administered the German version of the Structured Clinical Interview for Diagnostic and Statistical Manual (DSM)-IV diagnosis (First, Spitzer, Gibbon, & Williams, 1996). Patients were offered participation in the study, if Copyright © 2015 John Wiley & Sons, Ltd.
219 they met the following criteria: (a) DSM-IV (APA, 1994) criteria for major depressive disorder, recurrent, in partial remission (DSM-IV 296.35), and major depressive disorder, single episode, in partial remission (DSM-IV 296.25); (b) an initial score of ≥9 on the BDI-II (Beck, Steer, & Brown, 1996); (c) depression considered to be the most severe disorder if other comorbid disorders were present (i.e., severity status was the highest for the depressed symptomatology and participants reported to feel impaired mostly by their depression); (d) 18–65 years of age; (e) not meeting DSM-IV criteria for psychosis, mania, and current substance abuse and dependence; and (f) no other psychological treatment for the duration of the study. All questionnaires were presented in a paper–pencil form. Participants were randomly assigned to treatment (n = 31) or to a wait-list condition (n = 29). Patients on the wait-list received no treatment for at least 3 months. Since BADS data were missing from one person, sample size in the WAIT condition was reduced to n = 28. Cognitive-behavioral group treatment for depressive rumination(CBT-DR; Teismann et al., 2012) uses a combination of metacognitive therapy techniques, such as attention-training technique, disputation of positive and negative metacognitive beliefs, detached mindfulness (Wells, 2009) and strategies from behavioural activation, such as distraction, attending to sensory experiences, acceptance, problem-solving (Addis & Martell, 2004; Martell et al., 2001) to help patients overcoming depressive rumination. The treatment comprises 11 group sessions of 90 min. The treatment manual and the positive effects of the treatment on depression, rumination, metacognitive beliefs and depressive relapse have been described in detail elsewhere (Teismann et al., 2012, 2014). The main assessments reported in here were carried out at pretreatment/pre-wait-list (T1) and posttreatment/post-wait-list (T2).
Measures 1. BADS (Kanter et al., 2007; German version: Hoyer & Teismann, 2013). The BADS is a 25-item self-report measure with four subscales: Activation, Avoidance/Rumination, Work/School Impairment, and Social Impairment. Participants indicate to what extent each statement was true for them during the last week on a seven-point scale (not at all to completely). To score the BADS, items from all scales other than the Activation scale are reverse-coded, and then, all items are summed. To score the subscales, no items are reverse-coded. Adequate psychometric properties have been reported including good internal consistency, good test–retest reliability and evidence for construct validity for the total score and the subscale scores (Kanter et al., 2007, 2009). Clin. Psychol. Psychother. 23, 217–225 (2016)
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220 The German version of the BADS was developed by means of a translation–back-translation procedure according to relevant guidelines for the translation of psychometric instruments (Hambleton, 2001). First, the original English BADS was translated into German independently by the first and the last authors. Second, the two translated versions were translated back by two native speakers, and third, the two versions were evaluated, compared and approved—after minor wording adjustments. 2. BDI-II (Beck et al., 1996; German version: Hautzinger, Keller, & Kühner, 2006). The BDI is a 21-item selfreport measure that surveys depressive symptoms on a four-point severity scale. It has demonstrated good internal consistency (Cronbach’s α ≥ 0.89 across clinical and non-clinical samples) as well as adequate concurrent validity with clinical ratings of depression (Hautzinger et al., 2006). The internal consistency for the BDI-II was α = 0.87 in the SSA (SSA-I–SSA-III) and α = 0.89 in the clinical sample. 3. Beck Anxiety Inventory (BAI; Beck & Steer, 1993; German version: Margraf & Ehlers, 2007). The BAI is a 21-item questionnaire that measures cognitive and somatic symptoms of anxiety within the last week, with higher scores indicating increased anxiety (range = 0–63). Sample items include ‘unable to relax’ and ‘heart pounding or racing’. Good psychometric properties have been demonstrated among diverse samples (Beck & Steer, 1993). In the present study, internal consistency was α = 0.89 in the SSA (SSA-III) and α = 0.86 in the clinical sample. 4. Perseverative Thinking Questionnaire (PTQ; Ehring et al., 2011). The PTQ is a 15-item self-report measure that assesses characteristics of ruminative thinking (e. g., repetitive, difficult to disengage from, unproductive, capturing mental capacity). All items are answered on a five-point scale ranging from 0 (never) to 4 (almost always). The PTQ total score has an excellent internal consistency: Cronbach’s α = 0.95 (Ehring et al., 2011). Accordingly, internal consistency was good in the SSA (SSA-III) α = 0.95 and in the clinical sample α = 0.91. 5. Response Styles Questionnaire (RSQ; NolenHoeksema & Morrow, 1991; German version: Kühner, Huffziger, & Nolen-Hoeksema, 2007). The RSQ is a 32item instrument that was designed to measure the way in which an individual typically responds to feelings of depression or sad mood. Participants are asked to rate 21 ruminative responses (e.g., ‘I go away by myself and think about why I feel this way’) and 11 distracting responses (e.g., ‘I go to a favorite place to get my mind off my feelings’) to depressed mood on a four-point scale ranging from 1 (almost never) to 4 (almost always). The two subscales have been shown to have acceptable internal consistency (Kühner et al., Copyright © 2015 John Wiley & Sons, Ltd.
2007). Accordingly, internal consistency of the rumination subscale (SSA, SSA-III, α = 0.79; clinical sample, α = 88) and the distraction subscale (SSA, SSA-III, α = 0.84; clinical sample, α = 0.81) was good in the present sample. 6. Mindfulness Attention Awareness Scale (MAAS; Brown & Ryan, 2003; German version: Michalak, Heidenreich, Ströhle, & Nachtigall, 2008). The MAAS is a 15-item self-report instrument measuring the general tendency to be attentive to and aware of presentmoment experiences in daily life (e.g., ‘I find it difficult to stay focused on what’s happening in the present’; ‘I could be experiencing some emotion and not be conscious of it until sometime later’). It consists of a single factor and thus yields a single total score. In the present study, the MAAS has an internal consistency of α = 0.90 in the SSA (SSA-III) and of α = 0.89 in the clinical sample.
STATISTICAL ANALYSES SPSS 20 for Windows (IBM, Armonk, NY, USA) was used for the statistical analyses as well as AMOS 20 (IBM, Armonk, NY, USA), which was used to carry out confirmatory factor analysis. Internal consistencies of the derived scales were determined by calculating Cronbach’s α. Construct validity was evaluated for samples 1 and 2 by calculating Pearson correlation coefficients between the scales of the BADS and the other scales used in this study. Change sensitivity was assessed by analyzing differences between the treatment and the wait-list group at the post-treatment-wait assessment using one-way analyses of covariance, with pretreatment scores as covariate. Furthermore, significant pretreatment to post-treatment changes within the groups were analyzed by repeatedmeasure analyses of variances (ANOVAs). Within-group effect sizes (pre to post) were calculated according to the recommendation of Cohen (1988). All of the latter analyses were intent-to-treat analysis, with last data carried forward. Post data of two patients (6.5%) in the treatment group and four patients (14.3%) in the wait-list condition had to be inserted (cf. Teismann et al., 2014). Test–retest reliability, assessed with Pearson correlation coefficients, was established in participants of the wait-list condition, who returned for the post-wait assessment 3 to 5 months after the pre-wait assessment.
RESULTS Confirmatory Factor Analysis The hypothesized four-factor model of the German version of the BADS was investigated by confirmatory factor Clin. Psychol. Psychother. 23, 217–225 (2016)
Behavioral Activation for Depression Scale analysis using maximum-likelihood estimation and performing Bollen–Stine bootstrap because of nonnormality of the data. The sample consisted of n = 371 participants (n = 59 of them patients and n = 312 students). The overly sensitive chi-square (Tabachnick & Fidell, 2001) reached significance (χ 2 = 1031.875; df = 269; p < 0.001) indicating a poor model fit. Additional fit indices were regarded following the recommendations given by Hu and Bentler (1999). The root mean square error of approximation (RMSEA; which should be below 0.08) of 0.088, 90% confidence interval [0.082, 0.093] with p < 0.01, does not support an acceptable fit. The comparative fit index (CFI; which should be 0.95 or above) of 0.779 does not support model fit either. Only the standardized root mean residual (SRMR; a value of 0.11 or below indicating acceptable fit) of 0.0856 supports model fit. Therefore, modification indices were inspected. Following the highest modification index (MI) of MI = 128 864, fit indices were calculated taking into account the suggested
221 covariance between variables e11 and e12. Again, chisquare reached significance (χ 2 = 879 652; df = 268; p < 0.001). The RMSEA of 0.079 indicates moderate model fit as does the SRMR of 0.081. Still, the CFI of 0.822 does not indicate an adequate model fit. Figure 1 presents the completely standardized factor solution of the confirmatory factor analysis. The residuals (expressed as covariances) of the indicators were mostly in the small-to-medium range (0.00–0.78), suggesting that the model explained most indicators well. Most factor loadings are in an acceptable range (λ = 0.34–78) except for items 23 (λ = 0.16) and item 25 (λ = 0.17). The latent variable Activation is associated negatively with Avoidance/Rumination expressed by a covariance of 0.46, Work/School Impairment expressed by a covariance of 0.72 and Social Impairment expressed by a covariance of 0.31, while the latter three latent variables were associated positively (expressed by covariances ranging from 0.43 to 0.63).
Figure 1. Completely standardized factor solution of the confirmatory factor analysis. Act = Activation; Av_Ru = Avoidance/Rumination; Work_School_Imp = Work/School Impairment; Soc_Imp = Social Impairment; BADS = items of the Behavioral Activation for Depression scale
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Scale Properties Internal consistency was assessed using Cronbach’s α. The 25-item BADS had an overall α of 0.85 in both samples, respectively. The internal consistencies of the subscales were Activation (seven items, SSA: α = 0.76; clinical sample: α = 0.79); Avoidance/Rumination (eight items, SSA: α = 0.76; clinical sample: α = 0.76); Work/School Impairment (five items, SSA: α = 0.63; clinical sample: α = 0.60); and Social Impairment (five items, SSA: α = 0.81; clinical sample: α = 0.83). Having taken item number into account, all Cronbach’s α were sufficient (cf. Cortina, 1993). Test–retest reliability, assessed with Pearson correlation coefficients, was established on the 28 depressed participants of the wait-list condition (DOS), who returned for the post-wait assessment (T2) 3 to 5 months (M = 5.04, SD = 1.5 months) after the pre-wait assessment (T1). The BADS total scale (r = 0.51, p < 0.01), as well as the BADS subscales demonstrated sufficient test–retest reliability: Activation (r = 0.67, p < 0.001); Avoidance/Rumination (r = 0.62, p < 0.001); Work/School Impairment (r = 0.66, p < 0.001); and Social Impairment (r = 0.57, p < 0.01).
both samples. Only BADS Work/School Impairment was unrelated to depression severity in the DOS. Anxiety scores (BAI) were correlated with BADS scores in the SSA, but unrelated in the DOS. Regarding perseverative thinking scores (PTQ), correlations were in the expected direction in the SSA. In the DOS, perseverative thinking was strongly associated with the BADS Avoidance/ Rumination scale and moderately correlated with increased BADS Social Impairment scores. Depressive rumination (RSQ—Rumination Scale) was associated with BADS Avoidance/Rumination scores in both samples and with BADS Social Impairment in the DOS. Habitual distraction in response to depressed mood (RSQ— Distraction Scale) was associated with increased BADS Activation scores in both samples. Furthermore, increased distraction was associated with reduced Work/School Impairment in the DOS. Increased mindfulness (MAAS) was associated with reduced Avoidance/Rumination, Work/School Impairment and Social Impairment in the SSA, but was unrelated to BADS scores in the DOS.
Change Sensitivity Construct validity Table 1 presents the correlations between the BADS and its subscales on the one hand, and the criterion measures on the other hand. Regarding severity of depressive symptoms (BDI-II), correlations with the BADS total score and most subscales were in the expected direction—in Table 1.
In order to explore the ability of the BADS to track changes over the course of treatment, data steaming from the randomized controlled trial comparing CBT-DR with a wait-list control condition (Teismann et al., 2014) were analyzed. Table 2 shows participants’ scores at pretreatment wait and post-treatment wait as well as treatment effects. At pretreatment, ANOVA indicated that there were no
Correlations between BADS total/subscales and additional measures BDI
BAI
PTQ
RSQ-R
RSQ-D
MAAS
BADS total Student sample Outpatient sample
0.539** 0.429**
0.434** 0.220
0.466** 0.367**
0.231* 0.360**
0.214* 0.311*
0.376** 0.174
Activation Student sample Outpatient sample
0.246** 0.256*
0.261** 0.023
0.251* 0.068
0.132 0.161
0.307** 0.384**
0.110 0.064
Avoidance/Rumination Student sample Outpatient sample
0.512** 0.374**
0.472** 0.335*
0.472** 0.531**
0.216* 0.342**
0.057 0.080
0.424** 0.235
Work/School Impairment Student sample Outpatient sample
0.353** 0.182
0.236* 0.017
0.239* 0.104
0.190 0.207
0.151 0.316*
0.199* 0.108
Social Impairment Student sample Outpatient sample
0.429** 0.382**
0.183 0.201
0.312** 0.286*
0.102 0.303*
0.098 0.139
0.298** 0.204
BADS = Behavioral Activation for Depression Scale, BAI = Beck Anxiety Inventory; BDI = Beck Depression Inventory; MAAS = Mindfulness Attention and Awareness Scale; PTQ = Perseverative Thinking Scale; RSQ-D = Response Styles Questionnaire—Distraction Scale; RSQ-R = Response Styles Questionnaire—Rumination Scale. **p < 0.01, *p < 0.05
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Table 2. BADS outcome measures at each assessment CBT-DR (n = 31)
WAIT (n = 28)
Measure/time
M
SD
M
SD
Group effect†
BADS—total Pretreatment Post-treatment
87.45 96.61
21.81 18.84
83.71 84.78
16.79 20.37
F(1, 57) = 0.54 F(1, 56) = 5.69*
BADS—Activation Pretreatment Post-treatment
21.42 24.29
8.28 8.60
22.96 23.11
6.88 6.69
F(1, 57) = 0.59 F(1, 56) = 2.24
BADS—Avoidance/Rumination Pretreatment 22.16 Post-treatment 17.77
9.13 9.66
23.33 23.14
7.71 9.21
F(1, 57) = 0.27 F(1, 56) = 5.91*
BADS—Work/School Impairment Pretreatment 13.26 Post-treatment 11.09
5.88 7.19
10.96 11.28
5.95 6.09
F(1, 57) = 2.21 F(1, 56) = 2.84
BADS—Social Impairment Pretreatment Post-treatment
8.29 7.18
10.18 10.41
6.25 6.99
F(1, 57) = 0.09 F(1, 56) = 4.77*
10.77 7.29
†
At pretreatment, the group effect is based on a one-way analysis of variance. At post-treatment, the group effect is based on a one-way analysis of covariance, with pretreatment scores as the covariate. CBT-DR = cognitive-behavioral group treatment for depressive rumination; BADS = Behavioral Activation for Depression scale; M = mean; SD = standard deviation. *p < 0.05
significant differences between the groups on any BADS. At post-treatment, a series of one-way analyses of covariances indicated that CBT-DR was superior to WAIT regarding the BADS total score as well as the subscales Avoidance/ Rumination and Social Impairment. No group differences were observable on the BADS subscales Activity and School/Work Impairment. Repeated-measure ANOVAs were used to assess pretreatment-wait to post-treatment-wait change. CBTDR was associated with statistically significant changes of the BADS total [F(1, 30) = 13.11, p < 0.001; d = 0.47] as well as on all BADS subscales [BADS-Activity, F(1, 30) = 5.88, p < 0.05, d = 0.38; BADS-Avoidance/Rumination, F (1, 30) = 12.28, p < 0.001, d = 0.52; BADS-Work Impairment, F(1, 30) = 7.36, p < 0.01, d = 0.36; BADS-Social Impairment, F(1, 30) = 7.04, p < 0.01, d = 0.47]. No improvements at all were observed in the WAIT condition.
DISCUSSION In the present study, the factor structure, reliability, construct validity and change sensitivity of the German version of the BADS were investigated. Using confirmatory factor analysis, the four-factor structure reported by Kanter et al. (2007, 2009) for the original English version of the BADS provided a sufficient fit to the German version of the BADS. The original model was slightly adjusted following the highest modification index. Allowing for covariation of error variables e11 and e12 Copyright © 2015 John Wiley & Sons, Ltd.
resulted in a moderate-to-good model fit for the two fit indices (RMSEA and SRMR), while the CFI score was less impressive. With regard to contents, items 11 and 12 show a similarity that distinguishes them from the other indicators of the latent variable Activation. While all other five items refer directly to activation itself, items 11 (‘I did things even though they were hard because they fit in with long-term goals for myself’) and 12 (‘I did something that was hard to do but it was worth it’) additionally contain the denotation of bringing oneself to something even though it might be difficult (‘hard’). A model fit similar to our results has been described by Kanter et al. (2009) in their study on a community sample with elevated depression scores. Following their argumentation, it might as well be true for our results that the moderate sample size might have contributed to a fit less than perfect. In addition, Kanter et al. (2009) suggest considering investigating alternate model structures. The point that using a combined sample of a majority of students and a minority of patients for the confirmatory factor analysis might be responsible for a fit less than perfect can be ruled out; a tentative analysis only using the SSA provided equal results. Two items (items 23 and 25) of the BADS had quite low factor loadings. To ensure comparability with international studies, they were nonetheless retained in the questionnaire—a tentative evaluation omitting those items did not lead to any change in model fit. Internal consistencies and test–retest reliability were acceptable. (Only for one subscale the low number of items is presumably at the expense of comparably lower Clin. Psychol. Psychother. 23, 217–225 (2016)
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224 internal consistency.) This is especially interesting, since the test–retest interval in the current study was longer than in any study before and since BA is generally viewed as a state rather than a trait characteristic. Construct validity of the BADS was supported by expected associations between BADS total score and subscales and a set of relevant criterion measures of depression, perseverative thinking, depressive rumination, distraction and mindfulness. As in previous studies, depressive symptomatology was consistently negatively associated with BADS scores (Kanter et al., 2007, 2009; Raes et al., 2010). Associations with anxiety were less consistent (cf. Raes et al., 2010). As expected, it was found that the BADS-Avoidance/Rumination subscale correlated significantly with perseverative thinking and depressive rumination, respectively. Interestingly, there were also substantial associations between the BADS-Social Impairment subscale and perseverative thinking. This fits with earlier work showing that habitual rumination has a marked adverse effect on social relationships (e.g., Lam et al., 2003; Müller et al., 2012). Furthermore, it was shown that distractive responses to depressed mood, as assessed with the RSQ, are significantly correlated with increased activation, as assessed with the BADS-Activation subscale. Making use of distraction, in response to ruminative thinking as well as depressed mood, is actively trained in behavioural activation (Martell et al., 2001) and has repeatedly been shown to be an effective strategy for the short-termed regulation of emotional distress (Nolen-Hoeksema, Wisco, & Lyubomirsky, 2008). Taken together, most associations were stronger in the SSA than in the DOS. Therefore, future studies should reassess the clinical use of the BADS in larger samples of depressed patients. Different from what was expected, associations between mindfulness and BADS scores were only detected in the SSA but not in the outpatient sample. One possible explanation for this differential finding is that non-depressed students who engage in goal-relevant activities manage to do so in a mindful way, i.e., attentionally focused on what they are doing. Some (but not all) depressed outpatients, in contrast, may be more susceptible to internal and external distractions, and even though they become more active, they may have difficulties in focusing their attention on the task at hand. Finally, the BADS proved to be useful in tracking—even small to moderate—changes in activation reported by patients undergoing treatment. In general, CBT-DRs’ effect on facets of BA, such as an increase in Activity, and decreases in Avoidance behaviour (including Rumination) and Work and Social Impairment, go well with the therapeutic focus of the treatment. Although strategies like structured activity monitoring and scheduling were not an integral part of the treatment, patients were frequently reminded to become active when (and instead of) ruminating. Furthermore, strategies in dealing with rumination, such as distraction, attention to sensory experiences and Copyright © 2015 John Wiley & Sons, Ltd.
problem-solving (Martell et al., 2001) constituted a central part of the treatment. It is, however, unclear to what extent reductions in social impairments are due to these specific inputs of the therapy session. They may also have been instigated by the group setting itself: Thus, joining the group session itself may already significantly counteract social withdrawal behaviour. Furthermore, experiencing oneself in an interpersonal context may have helped to overcome feelings of social insufficiency and isolation. A set of further intriguing research questions would become testable by the use of the BADS in future studies on psychotherapy of depression: to what extent would different treatments for depression differ from each other in their effect on BA? do changes in activation predate and predict changes in depression? and finally, what is the importance of BA, as assessed with the BADS, for the prediction of depressive relapse and recurrences? Several limitations have to be considered, when interpreting the results. First, the sample size of the depressed outpatients was rather small. Future studies should attempt to utilize larger patient samples making it possible to evaluate the factor structure of the BADS separately for different clinical and non-clinical samples. Second, an acceptable model fit was only established after allowing for a—theoretically underpinned—covariation between two error variables. Therefore, a cross-validation of the here-described factor structure is necessary. In sum, the present results are largely consistent with those of previous research with the original English version of the BADS (Kanter et al., 2007, 2009). They extend previous research in several important ways by broadening information on reliability and construct validity (including theoretically relevant associations of the BADS, which had previously not been investigated) and employing a sample of depressed patients undergoing treatment.
ACKNOWLEDGEMENTS The study was funded by a grant from the German Research Society (Deutsche Forschungsgemeinschaft, DFG) to UW (WI 1106/12-1).
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