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The Relationship between Alliance & Outcome in PTSD
 

The Relationship between Alliance & Outcome in PTSD

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An important study about the relationship between alliance and outcome in the treatment of post traumatic stress disorder.

An important study about the relationship between alliance and outcome in the treatment of post traumatic stress disorder.

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    The Relationship between Alliance & Outcome in PTSD The Relationship between Alliance & Outcome in PTSD Document Transcript

    • Journal of Counseling Psychology Alliance and Outcome in Varying Imagery Procedures for PTSD: A Study of Within-Person Processes Asle Hoffart, Tuva Øktedalen, Tomas Formo Langkaas, and Bruce E. Wampold Online First Publication, August 19, 2013. doi: 10.1037/a0033604 CITATION Hoffart, A., Øktedalen, T., Formo Langkaas, T., & Wampold, B. E. (2013, August 19). Alliance and Outcome in Varying Imagery Procedures for PTSD: A Study of Within-Person Processes. Journal of Counseling Psychology. Advance online publication. doi: 10.1037/a0033604
    • Journal of Counseling Psychology 2013, Vol. 60, No. 4, 000 © 2013 American Psychological Association 0022-0167/13/$12.00 DOI: 10.1037/a0033604 Alliance and Outcome in Varying Imagery Procedures for PTSD: A Study of Within-Person Processes Asle Hoffart, Tuva Øktedalen, and Tomas Formo Langkaas Bruce E. Wampold University of Wisconsin Madison and Research Institute, Modum Bad, Vikersund, Norway This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. Research Institute, Modum Bad, Vikersund, Norway, and University of Oslo The present study examined both the intraindividual relationship between alliance components (task, goal, and bond) and subsequent posttraumatic stress disorder (PTSD) symptoms over the course of therapy and the interindividual relationships between the initial level of the alliance components and overall PTSD outcome. PTSD patients (n ϭ 65) were randomized to either standard prolonged exposure, which includes imaginal exposure (IE) to the traumatic memory, or modified prolonged exposure, where imagery rescripting (IR) of the memory replaced IE as the imagery component of prolonged exposure in a 10-week residential program. They were assessed repeatedly (weekly) on alliance and PTSD symptom measures. The centering method of detrending (Curran & Bauer, 2011) was used to separate the variance related to the intraindividual process of change during treatment (within-person component) from the variance related to initial individual differences (between-person component). The hypothesis of a negative within-person effect of the alliance components agreement about the tasks of therapy and bond on subsequent PTSD symptoms was supported for the component task agreement. As expected, this effect was stronger in IE than in IR. Moreover, there was a negative relationship between interindividual differences in initial Task and Bond scale scores and slope of PTSD symptoms over the course of therapy. By contrast, within-person variations in PTSD symptoms did not predict subsequent alliance components. The present results suggest the importance of agreement about therapy tasks during the process of IE or IR within prolonged exposure for PTSD patients, particularly in IE. Keywords: imaginal exposure, imagery rescripting, posttraumatic stress disorder, alliance, process research exposure (Foa, Hembree, & Rothbaum, 2007), which has been most extensively documented as an efficacious treatment for PTSD, consists of imaginal exposure (IE) to the traumatic memory, repeated listening to tapes of the imagery sessions, and in vivo exposure to avoided situations and stimuli. Thus, the patient is asked to approach what has evoked the most anxiety and distress. During IE, also the therapist is confronted with aversive information, which may evoke emotional responses he or she cannot express and consequently induce therapist feelings of powerlessness (Arntz, Tiesema, & Kindt, 2007). The strains put on the patient and the therapist potentially requires an agreement about these tasks, which suggests that both the patient and the therapist understand and accept the treatment rationale and believe that the treatment is an appropriate and beneficial approach to reduce symptoms (Keller, Zoellner, & Feeny, 2010). Furthermore, the traumatic experience and its aftermaths often involve helplessness, shame, guilt, and anger reactions that are difficult to reveal to another person (Lee, Scragg, & Turner, 2001). Many clients also fear that IE to the trauma experience will lead to loss of control and even insanity. All this requires a development of a bond, in which the patient trusts that the therapist understands, cares for, and accepts him/her and believes the therapist is able to help the patient regulate strong emotions (Wampold & Budge, 2012). Conversely, the demands of the trauma-focused procedures may lower many patients’ enthusiasm about engaging in the therapy A stronger therapeutic alliance has been found to be associated with better outcomes across a variety of treatment approaches and mental health problems (Flückiger, Del Re, Wampold, Symonds, & Horvath, 2012; Horvath, Del Re, Flückiger, & Symonds, 2011). According to the most widely accepted transtheoretical model, alliance is composed of agreement about the tasks of therapy, agreement about the goals of therapy, and the emotional bond between patient and therapist (Bordin, 1979). These components may have a different role and influence depending on the treatment approach and the problem being treated (Ulvenes et al., 2012; Webb et al., 2011). The task and bond components should be particularly influential on outcome in trauma-focused treatments of posttraumatic stress disorder (PTSD) because of the demands put on both the patient and the therapist. For instance, prolonged Asle Hoffart, Tuva Øktedalen, and Tomas Formo Langkaas, Research Institute, Modum Bad, Vikersund, Norway, and Department of Psychology, University of Oslo, Oslo, Norway; Bruce E. Wampold, Department of Counseling Psychology, University of Wisconsin Madison, and Research Institute, Modum Bad. Correspondence concerning this article should be addressed to Asle Hoffart, Research Institute, Modum Bad, N-3370 Vikersund, Norway. E-mail: asle.hoffart@modum-bad.no 1
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. 2 HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD tasks and also lead them to question the therapist’s care for them. Therefore, there should be much variation in the levels of task agreement and bond, both between patients and within patients over time, which in turn might well covary with outcome. An agreement about goals is necessary in trauma-focused therapy as well. However, there should be more uniformity of levels of agreement about the goals because the goals of trauma-focused therapy—reducing the fear of the trauma memory and of the trauma reminders (Foa et al., 2007)—should be strongly endorsed by the patients as well as the therapists. Consequently, restricted range reduces the possibility of finding covariation between agreements of goals and outcome. In general, task agreement has been found to be more strongly related to outcome than goal agreement and bond have (Horvath, 2011). With respect to PTSD patients, early alliance has been shown to predict their adherence to prolonged exposure (Keller et al., 2010) and their emotion regulation skills and outcome in a two-phase stabilization/skill development and exposure therapy for childhood abuse-related PTSD (Cloitre, Stovall-McClough, Miranda, & Chemtob, 2004). Based on this literature, we specifically expected that task agreement in particular, and perhaps bond as well, would predict better weekly as well as overall outcome in trauma-focused therapy of PTSD. Although the IE component of prolonged exposure is an effective intervention for trauma-related fear through the mechanisms of habituation and experienced nonoccurrence of feared event (Foa et al., 2007), it may be less effective for other trauma-related emotions such as shame, guilt, and anger. Repeated exposure to a traumatic memory involving shame and guilt may provide little corrective information and actually run the risk of reinforcing these emotions (Dalgleish & Power, 2004). To address the range of emotions in PTSD, some authors have advocated (Arntz et al., 2007) the addition of an element of imagery rescripting (IR; Smucker, 2005), in which an imagined change of the course of events of the trauma memory is induced. In a randomized controlled trial (RCT), Arntz et al. (2007) compared a combination of IE and IR to IE alone. They found no difference in reduction of PTSD severity but did find the IE and IR combination to be more effective for anger control, externalization of anger, hostility, and guilt, especially at 1-month follow-up. The IR method used in this study was to provide the patient with an opportunity to discover and express in imagery any trauma-related inhibited emotional responses (e.g., anger about what happened). The present study, the data for which was obtained in an RCT, replicates and extends the study of Arntz et al. by using a broader form of IR developed by Smucker (2005). In this method, the patient’s current self is—after an initial imagery reliving phase—invited to enter the imagery at the worst moment of the trauma, bring the situation to a solution (e.g., overpower a perpetrator), and then interact with the traumatized self back then. The patient’s anger is used as a resource in overpowering perpetrators and the current self– traumatized self interaction stimulates the development of selfcompassion instead of shame, guilt, and self-critique. The empowering and relieving features of IR may put less strain on the patient and the therapist by making them feel less helpless and distressed compared to IE and thus help them both to engage in imagery work. In the study by Arntz et al., therapists tended to favor the combination of IE and IR, as it decreased their feelings of helplessness compared to IE alone. Supporting the effectiveness of the broader form of IR, Grunert, Weis, Smucker, and Christianson (2007) found in an open trial that IR was extremely helpful for PTSD patients who had previously not profited from IE. The present study does not focus on therapy outcome per se but on how the influence of the alliance on outcome may relate to the specific trauma-focused therapy model being applied. We expected that, due to the empowering and relieving features of IR compared to IE, the influence of task agreement and bond on subsequent PTSD symptoms would be weaker in IR than in IE. Understanding the nature of the alliance depends on the methods used to examine it. For example, the well-established alliance/ outcome relationship is cross-sectional (i.e., bivariate observations for each psychotherapy dyad) and is thus focused on betweenperson differences (i.e., interindividual processes). That is, variations between patients in early alliance have been found to correlate with between-patient variations in outcome at the end of therapy (Horvath et al., 2011). However, it is also important to consider the development of the alliance for a particular patient. For example, the rupture-repair model (Safran & Muran, 1996) assumes that alliance ruptures represent opportunities for patients to learn about their problems relating to others, and repairs represent such opportunities having been taken in the here-and-now of the therapeutic relationship. This process is indicated by marked drops in alliance followed by a quick return to previous or higher levels, which represents within-person variations in the alliance. In general, therapy models, and particularly therapists, focus on within-person relationships, which would be the case, for example, when a change in the alliance for a particular patient leads to a subsequent alleviation of PTSD symptoms in that patient. The typical alliance data, collected once early in therapy, or occasionally during therapy, are unsuitable for evaluating withinperson processes (Curran & Bauer, 2011). Only repeated measures data allow for the proper disaggregation of between-person and within-person effects (Curran & Bauer, 2011; Hoffman & Stawski, 2009). Such a disaggregation not only allows the study of withinperson processes separated from between-person effects, but also is able to examine cross-level interactions of between- and withinperson effects. For instance, the effect of having a stronger alliance than expected for a particular patient may matter more for patients who have lower alliance in general. When the general (betweenperson) level of bond is low, for example when the patient has low trust that the therapist wants the best for him/her and is therefore preoccupied with this issue, a certain increase of this trust in a particular session might be a valued event with an immediate effect on symptoms. On the other hand, when the patient’s trust is already high and is not an issue for him/her, the same increase would probably have fewer consequences. That is, one should expect within-person variations in alliance to affect PTSD symptoms more when the between-person level of alliance is low. So far, the ability to separate these effects has not been fully capitalized upon in alliance research. Two notable exceptions are the studies of Tasca and Lampard (2012) and Falkenström, Granström, and Holmqvist (2013). Using latent change score modeling, in which between- and within-person components of both the predictor and outcome variables are separated, Tasca and Lampard obtained evidence for a reciprocal influence of alliance to the patient group and outcome among eating disordered individuals. Using the disaggregation methods in multilevel models proposed by Curran and Bauer (2011), Falkenström et al. also found evidence for a reciprocal causal model of alliance and outcome in
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. WITHIN-PERSON EFFECTS OF ALLIANCE ON OUTCOME primary care psychotherapy. Based on the results of these welldesigned studies, we expected to find that over the course of trauma-focused therapy, prior growth in task and bond would be associated with subsequent reduction in PTSD symptoms, and prior reduction in PTSD symptoms would be associated with subsequent growth in task and bond. The main purpose of the present study was to examine the role of alliance components in the process of therapeutic time-specific change in patients diagnosed with PTSD. The patients received either standard prolonged exposure, which includes IE, or modified prolonged exposure, where IR replaced IE as the imagery component of prolonged exposure, in a 10-week residential program. They were assessed repeatedly (weekly) on alliance and PTSD symptom measures, allowing us to separate the variance related to individual differences (between-person component) at the start of treatment from variance related to the intraindividual process of change during treatment (within-person component). To summarize, we wanted to examine the following hypotheses: Hypothesis 1: Time-specific change in a patient’s task and bond components of the alliance over the course of therapy are negatively related to subsequent change in PTSD symptoms assessed 3 days later (within-person effect). That is, when the task agreement and bond for a given patient is higher than is expected for that patient, subsequent symptoms will be lower. Hypothesis 2: Time-specific change in a patient’s PTSD symptoms over the course of therapy are negatively related to subsequent change in task agreement and bond assessed 4 days later (within-person effect). That is, when the PTSD symptoms for a given patient are less than is expected for that patient, subsequent task agreement and bond will be higher. Hypothesis 3: Individual differences in task agreement and bond at the start of imagery therapy are negatively related to individual differences in the rate of change of PTSD symptoms over the course of therapy (between-person effect). That is, patients who have a higher task agreement and bond at the start of imagery therapy will have a more negative rate of change of PTSD symptoms. Hypothesis 4: There is a cross-level interaction of betweenperson and within-person effects. That is, the lower the level of task agreement and bond is at the start of imagery therapy, the stronger the relationship between time-specific change in alliance and subsequent change in PTSD symptoms will be during therapy, and the higher the level of task agreement and bond is at the start of imagery therapy, the weaker the relationship between time-specific change in alliance and subsequent change in PTSD symptoms will be during therapy. Hypothesis 5: The within-person effect of task agreement and bond on subsequent PTSD symptoms is stronger for IE within prolonged exposure than for IR within prolonged exposure. We also wanted to explore the relationships between goal agreement and PTSD symptoms but expected the magnitude of this relation to be less than the magnitudes for task agreement and bond. 3 Method Participants The participants were selected from referrals to a PTSD treatment program at a national clinic. The clinic was established for the residential treatment of nonpsychotic patients who lack adequate local treatment opportunities or have not responded adequately to outpatient care and require more extensive and/or specialized treatment. The study eligibility was similar to treatment eligibility, that is, all patients who were considered to potentially benefit from the PTSD treatment were included. The inclusion criteria were (a) satisfying Diagnostic and Statistical Manual of Mental Disorders (4th ed.; DSM–IV; American Psychiatric Association, 1994) criteria for PTSD, (b) PTSD identified as the primary disorder in need of treatment, (c) age 18 to 67 years (regulated by the hospital), and (d) accepting withdrawal of all psychotropic medication (regulated by the hospital—patients referred to the hospital have usually received medication without effect). The exclusion criteria were (a) extensive dissociative symptoms, (b) current suicidal risk, (c) current psychosis, and (d) ongoing trauma (e.g., current involvement in an abusive relationship). The study was approved by the Regional Ethics Committee, and the patients’ gave informed consent after the procedures had been fully explained. A flow chart of patients is presented in Figure 1. Seventy-one patients were found eligible for treatment at the assessment stay and admitted to treatment from December 2008 to November 2010. At admission, all these 71 patients were found to meet research criteria, but three of them declined participation. One patient dropped out from treatment before randomization because she changed her mind about receiving trauma-focused therapy. The remaining 67 patients were randomized, 33 to IE within prolonged exposure and 34 to IR within prolonged exposure. Two IE patients lost their eligibility after randomization— one was found to need an eating disorder focus to the exclusion of imagery work, and another was inadvertently treated by the IR protocol. Thus, our intent-to-treat (ITT) with imagery sample consisted of 65 patients—31 IE and 34 IR patients—who signed consent, were randomized to an imagery condition, and were not removed by the investigators. Of these, three patients— one IE and two IR patients— dropped out within 5 to 6 weeks into the program. The reasons for dropout were conflict with therapist in two cases and serious somatic illness in one case. One IR patient received a restricted dose of rescripting, as she insisted to focus on her relationship to her parents after three sessions in accordance with the IR manual. This left a completer sample of 61 patients—30 IE and 31 IR patients. The mean age of 65 patients—38 women and 27 men—was 45.2 years (SD ϭ 9.7). The mean length of time since the index trauma was 17.5 years (SD ϭ 13.3). The most prevalent index trauma, defined as the one experienced by the patient as currently most distressing or most frequently reexperienced or both, among the 38 women was nonsexual assault by a familiar person (n ϭ 12; 31.6%), sexual assault by a familiar person (n ϭ 9; 23.7%), and sexual assault by a stranger (n ϭ 8; 21.1%). Among the 27 men, war experience was most frequent (n ϭ 7; 25.9%), followed by assault by a familiar person (n ϭ 6; 22.2%) and accidents (n ϭ 4; 14.8%). Over half the index traumas were prolonged and/or re-
    • HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD 4 Assessed for eligibility (N = 71) Excluded (n = 4) ♦ Declined to participate (n = 3) ♦ Dropped out before randomization (n = 1) This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. Randomized (n = 67) Allocated to imagery exposure (n = 33) ♦ Received full intervention (n = 30) ♦ ♦ Allocated to imagery rescripting (n=34) ♦ Received full intervention (n = 31) Lost eligibility (n = 2) Dropped out after 6 weeks (n = 1) ♦ Analyzed (n = 31) ♦ Excluded because lost eligibility (n = 2) Figure 1. Changed focus after 6 weeks (n = 1) ♦ Dropped out after 5 weeks (n = 2) Analyzed (n = 34) ♦ Excluded from analysis (n = 0) Flow of patients through the study. peated events. Among the 65 patients, 40 (61.5%) had current major depression or dysthymia, 44 (67.7%) had panic disorder with or without agoraphobia or agoraphobia without a history of panic disorder, 39 (60.0%) social phobia, 16 (24.6%) obsessive– compulsive disorder (Axis I), 11 (16.9%) generalized anxiety disorder, 18 (27.7%) alcohol abuse/dependence, 11 (16.9%) avoidant personality disorder, 9 (13.9%) substance abuse/dependence, and 9 (13.9%) obsessive-compulsive personality disorder. No other diagnosis exceeded a proportion of 10% in the present sample. According to chi-square tests, there were no diagnostic differences between the patients in the two treatment conditions. Measures PTSD Symptom Scale–Interview (PSS-I). The PSS-I (Foa, Riggs, Dancu, & Rothbaum, 1993) is a semistructured interview consisting of 17 items corresponding to the DSM–IV PTSD symptoms. Both PTSD diagnosis and PTSD symptom severity are assessed. Items are rated on 0 –3 scales for combined frequency and severity in the past 2 weeks (0 ϭ not at all, 1 ϭ once per week or less/a little bit, 2 ϭ 2 to 4 times per week/somewhat, and 3 ϭ 5 or more times per week/very much). Symptom severity is determined by the sum of the 17 ratings. The PSS-I has demonstrated satisfactory internal consistency reliability (Cronbach’s ␣ ϭ .85), high interrater agreement (interclass correlation [ICC] ϭ .97), high 1-month test–retest reliability (r ϭ .80), good concurrent validity with other measures of psychopathology, and excellent convergent validity with the Structured Clinical Interview for DSM–III–R (SCID; Spitzer, Williams, Gibbon, & First, 1988), correctly identifying the PTSD status of 94% of the studied subjects (Foa et al., 1993). The PSS-I was translated into Norwegian (see later) and used as the primary outcome measure in this study. Ten pretreat- ment and 10 posttreatment PSS-I interviews were randomly selected from the total sample of interviews and scored independently. Interrater agreement for the PSS-I total score was evaluated by means of ICC (3, 1; Shrout & Fleiss, 1979), with a value of .91 at pretreatment and .95 at posttreatment. PTSD Symptom Scale–Self-Report (PSS-SR). The PSS-SR (Foa et al., 1993) is a self-report version of the PSS-I and was used as a suboutcome measure in the present study. This measure is usually rated for the last week, but the rating period was shortened to the last 3 days in this study. The frequency part of the criteria was changed correspondingly (0 ϭ not at all, 1 ϭ 1 time/sometimes, 2 ϭ 2 times/half of the time, 3 ϭ 3 or more times/almost always. As for the PSS-I, symptom severity is determined by the sum of the 17 ratings. PSS-SR symptom severity has demonstrated satisfactory internal consistency reliability (Cronbach’s ␣ ϭ .91), high 1-month test–retest reliability (r ϭ .74), good concurrent validity with other measures of psychopathology, and excellent convergent validity with the SCID, correctly identifying the PTSD status of 86% of the studied subjects (Foa et al., 1993). The PSS-I and the PSS-SR were translated to Norwegian by the first and the third author and back-translated to English by a native-Englishspeaking professional also competent in Norwegian, until satisfactory formulations were found. Internal consistency reliability of the first-week PSS-SR rating was .88. One-week test–retest reliability coefficient for the PSS-SR scores from the first to the second week (before the more active therapy components were introduced) was .70. Concurrent validity was supported by a correlation of .68 between the first-week PSS-SR scores and pretreatment PSS-I scores. Working Alliance Inventory–Short Revised (WAI-SR). The WAI-SR (Hatcher & Gillaspy, 2006) is a shortened 12-item
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. WITHIN-PERSON EFFECTS OF ALLIANCE ON OUTCOME version of the original 36-item WAI (Horvath & Greenberg, 1989). Items are rated on a 1–7 Likert-type scale, and subscale scores for the task (four items), goal (four items) and bond (four items) components of alliance are computed by averaging across items. The WAI-SR has been found to differentiate well between these three components and has shown high internal consistency reliability (subscale score alphas ranging from .85 to .90) and high correlations with other alliance scales (Hatcher & Gillaspy, 2006). The WAI-SR has been translated to Norwegian and backtranslated to English until satisfactory formulations have been found (Horvath, 1981, 1984, 1991/2006). The internal consistency reliabilities of the four-item Task, Goal, and Bond subscales at the first assessment for the second week were .90, .91, and .85, respectively, and their 1-week test–retest reliabilities from the second to the third week were .72, .80, and .80, respectively. Procedure During a 3-day assessment stay, one of two research psychologists (the second and the third authors) evaluated the applicants by conducting the PSS-I to ascertain the diagnosis of PTSD, whereas the two individual therapists associated with the program evaluated the overall eligibility for the program. At the patients’ admission to the program (pretreatment), one of the two research psychologists conducted a comprehensive interview consisting of the PSS-I, the Mini International Neuropsychiatric Interview (MINI; Sheehan et al., 1994), and the Structural Clinical Interview for Axis II Personality Disorders (SCID-II; First, Spitzer, Gibbon, Williams, & Benjamin, 1994). The PSS-I was also conducted at discharge (posttreatment), but this time by a psychologist not involved in the study and blind to the patients’ treatment condition. The alliance measure (together with other process measures not analyzed here) was completed every Friday morning. The patients were asked to base their ratings on their experiences during the last 4 days, that is, during the most treatment-intensive part of the week. The PSS-SR was completed every Monday morning. The patients were asked to base their ratings on their experiences during the last 3 days, that is, during a less treatment-intensive period. To control for potential expectancy bias with respect to the alliance measure, patients were informed that the therapists were blind to the process ratings. Design and Randomization The patients received 10 individual sessions lasting 90 min over a period of 10 weeks. After 1 week of treatment (two first sessions according to the prolonged exposure protocol), the patients were randomized to either IE or IR as the imagery component of the treatment. A person who was not affiliated with the research team organized the randomization procedure. Random sequences generated from http://www.random.org were used for assignment to conditions. A blocked randomization procedure was used in which each therapist was assigned an equal number of cases in each condition. The probability of every patient ending up in any of the two conditions was kept constant at 0.5, and no measures were taken to correct for any imbalance in numbers between the conditions due to discontinued treatments. 5 Treatment The outpatient manuals for prolonged exposure, including IE (Foa et al., 2007) and IR (Smucker, 2005), were used but adapted for the inpatient setting. Essentially, it meant that milieu therapists were available to assist in between-session assignments (in vivo exposure, listening to tapes of the imagery work) and to provide safety and support after intensive individual sessions. The first two individual sessions were the same for all patients and consisted of giving a general treatment rationale and providing trauma education (first session) and introducing and planning in vivo exposure by constructing an exposure hierarchy (second session). Then, before the third session, patients were stratified by therapist and randomly allocated to either the IE or the IR condition, after which they followed the relevant protocols for the third (occurring toward the end of the second week of treatment) to ninth session. In the tenth and final session, the content was again identical and consisted of imagery exposure to the total memory, a review of progress, and suggestions of continued practice. In the sixth week, the patients returned home to test their newly acquired skill in their natural environment. All the time, there was one other treatment group of anxiety patients at the ward, and the PTSD patients participated in the ward’s general program, consisting of one physical exercise session and one ward meeting per week. The IE approach consisted of having participants relive the traumatic event in their imagination and recount the memory in the present tense. To increase vividness, patients were asked to report as much detail as possible, including sights, sounds, smells, behaviors, bodily sensations, feelings, and thoughts. The memory was repeated if necessary to allow total reliving for a period of 40 to 60 min. The entire memory was relived during the first two or three sessions. In the subsequent sessions, the hot spots procedure was usually applied, where reliving was focused on the currently most distressing parts of the memory. The IR approach consisted of three continuous phases. The first phase consisted of imagery reliving of traumatic event in order to activate the trauma memory and to identify the hot spot(s). In Phase 2, without pause in imagery, the memory was relived from the beginning, but this time—at the identified hot spot—the patient was asked to imagine the current self entering the scene at the hot spot and bringing the situation to a solution (overpowering the perpetrators or updating the traumatized self back then with future information). Finally, in Phase 3, patients were stimulated to imagine an interaction between the current self and the traumatized self back then. As in IE, the imagery was supposed to last 40 to 60 min. Therapists One of the individual therapists was a 57-year-old male clinical psychologist with a PhD. The other was a 55-year-old female psychiatric nurse with a master’s degree. The milieu therapists were four psychiatric nurses ranging from 45 to 60 years old. All the individual and milieu therapists had at least 10 years of experience in the cognitive therapy programs for anxiety disorders at the unit and had completed the cognitive therapy specialization program provided by the Norwegian Association of Cognitive Therapy. Of the 65 ITT patients, the psychologist treated 16 IR patients and 15 IE patients, whereas the nurse individual therapist treated 18 IR patients and 16 IE patients.
    • 6 HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD Training and Supervision All the staff received prestudy workshops and supervision by experts Elizabeth Hembree (in prolonged exposure including IE) and Mervin R. Smucker (in IR) during several pilot treatment groups. Throughout the study period, all of the individual sessions were videotaped, and each of the experts provided 90-min supervision sessions of taped imagery biweekly. In addition, the first author provided two 60-min supervision sessions per week to the milieu staff and individual therapists in a group format. This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. Treatment Integrity The Treatment Integrity Checklist (Foa, Hearst-Ikeda, Dancu, Hembree, & Jaycox, 1997) contains items describing essential and desirable ingredients of prolonged exposure therapy across the 10 sessions described in the manual (Foa et al., 2007). As we were particularly interested in assessing the imagery component, which was the only intended difference between the two treatment conditions, we rated the eight items of the Prolonged Exposure Sessions 4 –9, Section C: Imaginal Exposure. Three of the eight items refer to ingredients that are obligatory (e.g., “reviews instructions for imaginal exposure”), whereas five refer to ingredients that should be present if needed (e.g., “titrates the experience as needed”). Based on discussions with the originator of IR, Mervin R. Smucker, a corresponding checklist for this method was constructed. It consisted of the same three obligatory items as for the IE checklist, one unique obligatory item, and six unique per-asneeded items (e.g., “identifies relevant action impulses coming from the client and helps the client to implement them within imagery”). A score for percentage adherence during an imagery episode is computed by dividing the number of obligatory and per-as-needed ingredients present by the total number of items rated. An overall adequacy (competence) rating for the episode was given using a 1–5 scale with the anchor points poor, mediocre, satisfactory, good, and excellent. Finally, the presence or absence of IR elements was rated. The expert on the therapy form (Elizabeth Hembree or Mervin R. Smucker) rated the episode together with the first and the second author, whereas the third author did simultaneous translation of the videotape. A pilot case in each therapy form was first rated and discussed to calibrate the ratings. Then, 10 random cases, stratified for order of treatment group in the trial and individual therapist, from each therapy form were selected. From these 20 cases, the imagery part of the fifth individual session was rated. One of the cases turned out to be the one who was inadvertently treated by IR instead of IE (see Participants), and this case was omitted from all analyses. Thus, 19 (4.3%) of the total of 440 sessions including the specific imagery component were analyzed. The intraclass correlation (ICC [3, 2]; Shrout & Fleiss, 1979) was .69 in IE and .92 in IR for adherence and .93 in IE and .87 in IR for adequacy. The results are based on the expert ratings. Mean adherence rating was 75% (SD ϭ 15%) in IE and 80% (SD ϭ 21%) in IR. Mean adequacy rating was 2.78 (SD ϭ 1.30) in IE, corresponding to a level a little below satisfactory, and 3.20 (SD ϭ 1.32) in IR, corresponding to a level a little above satisfactory. One minor protocol violation was detected in one of the IE sessions, where the therapist asked questions typical of IR for a couple of minutes. After the trial, we asked the individual therapists to fill in a questionnaire about their preference for IE or IR. The psychologist indicated no preference, whereas the psychiatric nurse reported preference for IR because she felt patients’ experience of taking the power from the perpetrator was particularly helpful. Statistical Analysis A main purpose of this study was to examine how within-person changes in components of alliance affected subsequent withinperson changes in outcome. Such a focus on within-person processes necessitates a proper disaggregation of the within-person and between-person components of change in the time-varying predictor. The choice of method of disaggregating within-person and between-person effects in a time-varying predictor depends on how it is related to time (Curran & Bauer, 2011). Specifically, it is important to know if this relationship is characterized by a fixed effect of time or if it is characterized by both a fixed and random effect of time. To estimate these parameters, we conducted several series of mixed models using the three alliance scales (WAI-Task, WAI-Goal, WAI-Bond) and the PTSD symptom measure (PSSSR) as dependent variables. The intent-to-treat sample was analyzed, and due to our research purposes, scores were included from the start of the imagery part of therapy (from the second week of treatment). Moreover, as only active treatment time was of interest, ratings from the week at home were not included, and the home week was not counted in the time term. The fit of these nested models for the covariance was compared by using the likelihood ratio test, in which the difference in model –2 log likelihood values is divided by the difference in degrees of freedom of the models (Fitzmaurice, Laird, & Ware, 2004). Restricted maximum likelihood estimation was used to estimate nested models with only varying random effects (Fitzmaurice et al., 2004). Models with different fixed effects were compared using maximum likelihood estimation. We used an unstructured covariance structure for the random effects, thus allowing the estimation of covariance between the random intercepts and slopes. By contrast, we used a diagonal covariance structure for the residuals, thus allowing the variances of the residuals to differ over time points but setting the covariance between the residuals across time points to zero. Thus, the correlation between the scores across assessments had to be modeled exclusively by the random effects. We started with a model with only a fixed intercept and no random effects, added a random intercept, and, finally, added a random effect of week in therapy. After the best random effects structure had been found in this way, we tested whether another residual covariance structure besides the diagonal—for example, a first-order autoregressive (e.g., AR(1), Toeplitz)— could improve model fit. We then tested whether the inclusion of a fixed linear time term (week in therapy) and—in a second step—a fixed quadratic time term (week2) as independent variables improved model fit. Again, the fit of these nested models was compared by using the likelihood ratio test. For all the alliance scales, a fixed and random intercept and a fixed and random linear effect of time gave the best model fit. Moreover, no alternative residual covariance structure to the diagonal turned out to improve the fit. For the PSS-SR scores as well, a fixed and random intercept and a fixed and random effect of time turned out to be the most appropriate model. In addition, an AR(1) residual covariance structure improved model fit compared to the diagonal structure.
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. WITHIN-PERSON EFFECTS OF ALLIANCE ON OUTCOME In order to disaggregate the within-person and between-person variability in the time-varying alliance and symptom measures, we utilized the statistical centering method of detrending presented by Curran and Bauer (2011). This method was chosen because these time-varying predictors were characterized by a fixed and random effect of time. We created two new variables representing the within-person change and between-person differences for bond, task, goal, and PSS-SR scores, respectively (see the applied equations in the Appendix). First, we created the within-person predictor by regressing the variables on time separately for each individual using ordinary least squares. The resulting within-person deviations over weeks in therapy represent the within-person component of the time-varying alliance and symptom measures. In this way, the within-person deviations are conceptualized as the difference between a time-specific observation and the trend line for the variable (i.e., the expected value given a linear growth in the variable). Due to our present research purpose to examine the effect of between-person differences in alliance at the start of the differing imagery therapies, we used the estimated differences on the timevarying predictors at this time point (second week of treatment) to represent their between-person component. By setting time to zero at this point, the between-person component of the time-varying measures are represented by the estimated intercept at the second week for each individual. To correct for the possibility of Type I error, the chosen alpha significance level of .05 was divided by the number of tests (two) for each hypothesis, yielding a level of .025 for the individual test. Because all of the hypotheses were directional in nature, one-tailed tests were used. The effect size (ES) of the overall outcome was computed as Hedges’s g for dependent samples (Borenstein, Hedges, Higgins, & Rothstein, 2009). ESs of the between-person and within-person effects were calculated as the proportion of explained outcome variance for each predictor (Snijder & Bosker, 1999; see the applied equations in the Appendix). We used the program SPSS 19.0. Results Overall Outcome In the following ITT analyses, pretreatment PSS-I ratings substituted missing posttreatment ratings. Due to a failure in administrative routines, one IE patient missed the pretreatment PSS-I interview, and his ratings were substituted by the first and the last PSS-SR score. On the PSS-I, the 34 IR patients changed from 33.32 (SD ϭ 6.88) at pretreatment to 22.71 (SD ϭ 14.27) at posttreatment, yielding an ES of Ϫ0.83, 95% CI [Ϫ0.46, Ϫ1.20]. The corresponding change among the 31 IE patients was from 35.19 (SD ϭ 8.24) to 19.90 (SD ϭ 13.76), with an ES of Ϫ1.27, 95% CI [Ϫ0.76, Ϫ1.78]. In the total sample of 65 patients, the ES was Ϫ1.06, 95% CI [Ϫ0.74, Ϫ1.38]. A time by treatment repeated-measures analysis of variance yielded a time effect, F(1, 63) ϭ 69.87, p Ͻ .0001, but no treatment effect, F(1, 63) ϭ 0.04, ns., or time by treatment effect, F(1, 63) ϭ 2.27, p ϭ .137 (two-tailed). 7 Summary Statistics for the Weekly Outcome and Alliance Measures Missing data in the intent-to-treat sample during active imagery treatment was 6.4% for PSS-SR scores, 9.8% for Task scores, 10.0% for Goal scores, and 10.5% for Bond scores. The mean between-person PSS-SR score at the second week (estimated intercept) was 31.31 (SD ϭ 9.06). At the second week of treatment, mean between-person Task score (estimated intercept) was 5.33 (SD ϭ 1.17), Goal score was 5.65 (SD ϭ 1.12), and Bond score was 5.14 (SD ϭ 1.34). An F test for comparing variances in correlated variables showed that the standard deviations of the between-person Task, Goal, and Bond scores were not significantly different. The standard deviations of the within-person Task, Goal, and Bond scores were 0.4752, 0.4367, and 0.4127, respectively. An F test showed that Task scores had larger variances than did Goal and Bond scores (both ps Ͻ .025). The intercorrelations for the estimated between-person alliance scores at the second week (intercept) were high: .87 for Task and Goal, .62 for Task and Bond, and .73 for Goal and Bond. The intercorrelations for the within-person alliance scores over the course of imagery treatment were more moderate: .64 for Task and Goal, .46 for Task and Bond, and .51 for Goal and Bond. Testing Hypotheses Our weekly outcome measure—the PSS-SR—was used as dependent variable in mixed models with random intercept and slope and an AR(1) covariance structure for the residuals (see the Statistical Analysis section). Time (week), treatment (IR vs. IE), and the within-person and between-person components of the three WAI scales were used as predictors. Separate analyses were conducted for each scale. To establish a temporal sequence between predictor and outcome, within-person alliance scores were lagged and thus related to the PSS-SR scores the following week (3 days later). A summary of the fixed main effects for the three alliance components (viz., task, goal, and bond) on PTSD symptoms, as well as the random effects, are shown in Table 1. Our first hypothesis, about a negative within-person effect of task agreement and bond on subsequent symptoms, was supported for the Task scale. That is, if a patient had stronger agreement on tasks in a given week than would be predicted for that patient given his/her general trend, then this patient’s subsequent (3 days later) symptoms were lower than would be expected. The Goal and Bond scales showed no such within-person effect. Unrelated to our hypotheses, Table 1 also shows that there was a negative relationship between interindividual differences in initial Task scores and mean level of PTSD symptoms over the course of therapy but no such relationship for the other two WAI scales. In addition, there was a negative effect of time, which indicates that the PSS-SR scores were reduced over the course of therapy. There was no effect of treatment (viz., IR vs. IE) on the mean level of PSS-SR scores over the course of therapy. To examine our hypothesis about reciprocal causation, that is, that the PTSD symptoms would be negatively related to subsequent task agreement and bond, the three WAI scales were used as dependent variables in mixed models with random intercept and slope and a diagonal covariance structure for the residuals (see the Statistical
    • HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD 8 Table 1 Fixed Effects Estimates and Random Effects (Variance–Covariance) Estimates for the Three Models of the Predictors of PTSD Symptoms This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. Parameter Intercept Week Treatment: IR Treatment: IE Within-person task Between-person task Within-person goal Between-person goal Within-person bond Between-person bond Residual AR(1) rho Intercept Week Intercept ϫ Week –2 log likelihood Task Goal Bond Fixed effects 40.885‫)602.6( ء‬ 43.138‫)453.5( ء‬ Ϫ1.312‫)071.0( ء‬ Ϫ1.272‫)361.0( ء‬ Ϫ2.446 (2.265) Ϫ3.470 (2.374) 0 (0) 0 (0) Ϫ0.820‫)193.0( ء‬ Ϫ2.139‫)779.0( ء‬ Ϫ0.527 (0.426) Ϫ1.667 (1.065) 37.054‫)179.4( ء‬ Ϫ1.316‫)771.0( ء‬ Ϫ1.110 (2.411) 0 (0) Ϫ0.724 (0.464) Ϫ1.174 (0.910) Random effects 17.136‫)249.1( ء‬ 16.638‫)997.1( ء‬ 0.210‫)980.0( ء‬ 0.193‫)680.0( ء‬ 68.460‫)872.41( ء‬ 79.295‫)442.61( ء‬ 1.118‫)003.0( ء‬ 1.235‫)433.0( ء‬ 2.947 (1.617) 3.897‫)856.1( ء‬ 2737.418 2727.652 16.096‫)348.1( ء‬ 0.191‫)390.0( ء‬ 74.918‫)367.51( ء‬ 1.296‫)343.0( ء‬ 2.712 (1.752) 2549.456 Note. Standard errors are in parentheses. PTSD ϭ posttraumatic stress disorder; IR ϭ imagery rescripting; IE ϭ imaginal exposure; task ϭ agreement about tasks; goal ϭ agreement about goals; bond ϭ patient–therapist emotional bond; AR(1) ϭ first order autoregressive. ‫ء‬ p Ͻ .05. Analysis section). Within-person and between-person PSS-SR scores were used as predictors. In addition, we included time and treatment as predictors of the alliance scores. Our hypothesis that within-person variations in PTSD symptoms would predict subsequent withinperson variations in PTSD symptoms was not supported for any of the WAI scales. That is, there was no within-person effect of PSS-SR scores on Task, Goal, or Bond scores (all absolute t values Ͻ 1). To examine our third to fifth hypotheses, all the six interactions between our four predictors were added in the three models. Our third hypothesis, stating that higher initial task agreement and bond predicted a steeper negative slope of PTSD symptoms, was supported. That is, there was a significant time by between-person task effect, ␤ ϭ Ϫ0.272, SE ϭ 0.136, t(55.5) ϭ Ϫ2.00, p ϭ .025, and a significant time by between-person bond effect, ␤ ϭ Ϫ0.337, SE ϭ 0.125, t(55.6) ϭ Ϫ2.71 p Ͻ .01. As these interaction effects were negative, they indicate that with longer time into therapy, higher initial alliance was associated with lower PTSD symptoms. There was no time by between-person goal effect on symptoms. Our fourth hypothesis, that the within-person effect of alliance on outcome is stronger with lower initial levels of alliance, was contradicted by the results for the Task scale. That is, there was a cross-level interaction of between-person and within-person effects of task, ␤ ϭ Ϫ0.814, SE ϭ 0.403, t(320.7) ϭ Ϫ2.02, p Ͻ .025. The negative direction of this interaction effect shows that— opposite to what we expected—the higher the initial task alliance, the stronger the negative relationship between within-person variations in task alliance and subsequent within-person variations in PTSD symptoms. No crosslevel interactions of the within- and between-person effects were evident for the Goal and Bond scales. Our fifth hypothesis, stating that the within-person relationship between alliance and outcome is stronger in IE than in IR, was supported for the Task scale. That is, treatment interacted with the within-person effect of Task scores on PSS-SR scores. When using IE as a baseline, there was a positive effect of IR on PSS-SR scores, ␤ ϭ 2.031, SE ϭ 0.775, t(325.4) ϭ 2.62, p Ͻ .01. Considering the overall negative within-person effect of Task scores on PSS-SR scores (see Table 1), the positive direction of this relationship in IR compared to IE shows that the relationship is weaker in IR than in IE. It should also be noted that there was a time by treatment effect. In the model using task as a predictor, there was a positive effect of IR on PSS-SR scores with time, ␤ ϭ 0.964, SE ϭ 0.317, t(54.9) ϭ 3.04, p Ͻ .01. Considering the overall negative effect of time on PSS-SR scores (see Table 1), the positive effect of IR compared to IE used as baseline shows that the PSS-SR scores were less reduced in IR than in IE. There was no individual therapist effect on the rate of change of PSS-SR, Task, Goal, or Bond scores. The Magnitude of Effects Compared to a baseline model including only the random effects (intercept, time) and the fixed effect of time, residual variance was reduced, with 4.3%, while random intercept variance was reduced, with 5.8%, when within-person and between-person Task scores were added in the model. Discussion The Role of Alliance in Varying Imagery Procedures for PTSD The main purpose of this study was to examine the role of alliance components in the process of therapeutic change in PTSD patients. Most importantly, the hypothesis of a negative within-
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. WITHIN-PERSON EFFECTS OF ALLIANCE ON OUTCOME person effect of the components agreement about the tasks of therapy and bond on subsequent PTSD symptoms was supported for the task component. That is, when the task score for a given patient was higher than was expected for that patient, the subsequent symptom score was lower than was expected for him/her, explaining about 4% of the outcome variance. In any event, this finding goes beyond previous research in the PTSD treatment field by indicating that time-specific change in a person’s task agreement during therapy is related to this person’s subsequent change in PTSD symptoms. On a more general level, this finding supports and extends those of Tasca and Lampard (2012) and Falkenström et al. (2013), who found a within-person relationship between overall alliance and subsequent symptoms but in different treatments and patient populations. Furthermore, the results indicate that the within-person relationship between task agreement and outcome is dependent on the specific therapy form. As we hypothesized, this relationship was stronger in IE than in IR. On the other hand, within-person changes in PTSD symptoms did not predict subsequent task agreement and bond. That is, time-specific change in a person’s symptoms during therapy was not related to this person’s subsequent change in task and bond. This finding is at odds with those of Tasca and Lampard (2012) and Falkenström et al. (2013), who found a bidirectional relationship between alliance and symptoms. A conspicuous difference between these studies and ours is that we used standardized, highly structured, and manual-based procedures. One may speculate that patients’ belief in and agreement to such procedures are less influenced by symptom variations than is their agreement to less standardized and less clearly defined procedures. Our hypotheses about a between-person effect of initial task agreement and bond was supported. Initial Task and Bond scores predicted a steeper negative slope of PTSD symptoms. These findings are consistent with most findings in alliance research (Horvath et al., 2011) that early alliance predicts the further course of symptoms. The centrality of the task component in predicting overall (between-person) outcome is consistent with the results of Webb et al. (2011), who found therapist–patient agreement on the tasks and goals of therapy to account for most of the outcome variance in cognitive therapy for depression. However, our results indicate that a good initial bond is also important for a successful overall outcome in exposure-based therapy for PTSD. Thus, alliance components may have different roles in cognitive behavioral therapy (CBT) for different patient populations. What results would be obtained for forms of therapy other than CBT is unknown, as it appears that the bond works differently in dynamic therapy than it does in CBT. For instance, the bond and therapist’s focus on affect seem to be differently related to each other and to outcome in these two therapies (Ulvenes et al., 2012). We expected that task agreement and bond would be of greater concern for those who had a lower individual level and would thus be more influential in these persons’ process of change, but in fact the within-person effect of task scores on subsequent PTSD symptoms was stronger in those with a higher initial task agreement. Future research must show whether this was a chance finding or not. However, if this effect is replicated, it would suggest a double drawback for a patient having a low initial agreement about the tasks of therapy. First, the patient would experience less overall improvement over the course of therapy, and, second, greater than 9 expected levels of agreement during the process of therapy for that patient would not be as effective. We also explored the role of the alliance component goal agreement. As expected, this component was unrelated to symptom change. The within-person component of goal agreement also had less variance than the within-person component of task agreement, and this difference may have contributed to the differential findings. Our results are consistent with Horvath (2011), who found— on the between-person level—that the agreement on tasks as a predictor of outcome was superior to both bond and agreement on goals. Strengths and Limitations of the Study Alliance and PTSD symptoms were assessed weekly, and adequate methods were utilized to separate the within-person and between-person effects of the time-varying predictors in the applied multilevel models. Thus, we could study within-person relationships over the course of therapy, which are of particular relevance for psychotherapy theories. This is because therapy theories concern such relationships, that is, how change in a process variable relates to subsequent change in an outcome variable. Such knowledge directly informs therapists concerning what process variables need to be affected to achieve patient improvement. By contrast, knowledge of between-person relationships— one patient having a low initial alliance and poor outcome and another having a high initial alliance and good outcome— does not imply that an increase in the first patient’s alliance would lead to a better outcome for that patient. Thus, relationships established on a between-person level do not imply that the same relationships hold on a within-person level. For instance, the relationship between bond and outcome obtained in the present study on the between-person level was not replicated on the within-person level. A further advantage of properly separating the between- and within-person components of a time-varying predictor is the possibility of examining cross-level interactions of within- and between-person effects. For therapists, how between-person differences in, for example, alliance or self-concept moderate withinperson relationships over the course of therapy is more directly relevant than are the correlations of these differences with overall outcome. Such moderating knowledge informs therapists concerning under what conditions (e.g., high task agreement relative to other patients) certain within-person change processes are working (e.g., higher than usual task agreement at a given time point predicts lower than usual PTSD symptoms). A further advantage of studying within-person relationships between process and outcome is the possibility of identifying reciprocal or even reversed causality between process and outcome. The RCT design, where patients were randomized to two empirically based imagery methods, allowed us to study the moderating influence of therapy form on the within-person relationships. The studied sample had high clinical representativeness, as research eligibility was similar to treatment eligibility and only three (4.2%) of 71 treatment eligible patients declined research participation. Moreover, the dropout rate from imagery treatment was low: three (4.6%) of 65 patients. The present study has several limitations. Although the uncontrolled effect size of Ϫ1.27 (Hedges’s g, intent-to-treat analysis) for standard prolonged exposure (including IE as the imagery component) is comparable to that in one of the studies conducted
    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. 10 HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD by the originators of prolonged exposure (e.g., Hedges’s g ϭ Ϫ1.37 in Foa et al., 2005), the adequacy ratings were only around a level of satisfactory for both imagery treatments. The adherence ratings of 75% (IE) and 80% (IR) are lower than those typically found in the original studies of prolonged exposure (e.g., 97% in Foa et al., 2005). Thus, the varied component of treatment may have been delivered in a less than optimal way. Moreover, less extensive examinations of integrity (4.3% of tapes) than usual (about 10% of tapes; Foa et al., 1997) were performed. No integrity ratings were performed for the other components of treatment (e.g., in vivo exposure). Although well-validated measures were used, their Norwegian translations have not undergone psychometric evaluation in previous investigations. In the present study, their internal consistency and test–retest reliability appeared satisfactory, though. Alliance and symptom ratings were collected from the same individual, that is, the patient, and this may have inflated their correlation. However, halo effects were prevented by having the ratings done 3 and 4 days apart. Furthermore, response biases like acquiescence are supposed to cut across ratings and may affect within-person variations—which were the main focus of this study—to a lesser degree. We used a passive observational design, and unmeasured third variable confounds could have influenced the results. The power of the study, based on about eight repeated measurements of 65 patients (minus some missing data), may be too low to detect some withinperson relationships. We studied process on a weekly time scale, and larger or lesser scales could be associated with different results. The strategy of using the same therapists across therapies has both strengths and weaknesses. The therapists may not be equally competent and have the same preferences for both therapies. Actually, one of the therapists reported a preference for IR. However, this bias could not explain the present results, as PTSD symptoms measured weekly were less reduced over the course of therapy in IR than in IE. In the context of the present study, an advantage of crossing therapists was that the general ability to form alliances was balanced between conditions. Research Implications As elaborated above, our study invites an increased focus on within-person relationships in psychotherapy research. In highly structured therapies like those of the present study and cognitive therapy of depression (Webb et al., 2011), symptomatic improvement is supposed to result from the relatively specific tasks of these therapies. Agreement about tasks may therefore be particularly important in such therapies. Moreover, the studied PTSD sample was a severe one with a high degree of comorbidity and a long duration of PTSD, and over half of the patients had experienced repeated and/or prolonged traumas. Future studies should investigate the within-person relationships between alliance components and outcome across therapies and type and severity of disorders. Furthermore, studies of within-person relationships between therapy events/therapist actions and alliance components are needed. Clinical Implications The present within-person results make a firm basis for the recommendation to monitor, increase, and restore decreases of agreement about therapy tasks over the course of IE or IR within prolonged exposure for PTSD patients. They also suggest that addressing agreement about the tasks of therapy is particularly important in IE compared to IR. Given that these exposure methods consist of confronting the feared trauma memory and feared external situations, agreeing to their use based on an understanding of the rationale for these methods and a belief in their efficacy seems paramount. On the other hand, the results do not imply an increased focus on the agreement about goals of therapy and bond components of alliance over the course of these treatments. Our between-person results may inform therapists using prolonged exposure for PTSD that low initial task agreement and bond signal a poorer outcome of therapy. 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    • This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. WITHIN-PERSON EFFECTS OF ALLIANCE ON OUTCOME Foa, E. B., Hembree, E., Cahill, S. E., Rauch, S. A. M., Riggs, D. S., Feeney, N. C., & Yadin, E. (2005). Randomized trial of prolonged exposure for posttraumatic stress disorder with and without cognitive restructuring: Outcome of academic and community clinics. Journal of Consulting and Clinical Psychology, 73, 953–964. doi:10.1037/0022006X.73.5.953 Foa, E. B., Hembree, E., & Rothbaum, B. O. (2007). Prolonged exposure therapy for PTSD: Emotional processing of traumatic experiences. New York, NY: Oxford University Press. Foa, E. B., Riggs, D. S., Dancu, C. V., & Rothbaum, B. O. (1993). Reliability and validity of a brief instrument for assessing post-traumatic stress disorder. 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    • HOFFART, ØKTEDALEN, LANGKAAS, AND WAMPOLD 12 Appendix Equations Used in the Statistical Analyses ␤1i ϭ ␥10 ϩ u1i Equations for the Multilevel Models We begin with the Level 1 model: yti ϭ ␤0i ϩ ␤1i x ti ϩ eti u0i ϭ (zi Ϫ ␥00) Ϫ (␥10 ϩ u1i)xi ៮ (A1) Composite: This document is copyrighted by the American Psychological Association or one of its allied publishers. This article is intended solely for the personal use of the individual user and is not to be disseminated broadly. eti ϭ zti Ϫ ␤0i Ϫ ␤1i x ti . Ei(zti) ϭ (␥00 ϩ u0i) ϩ ␥10Ei(x ti) ϩ Ei(u1i x 0i) Here the individual-specific value of posttraumatic stress disorder (PTSD) symptoms (yti) is a function of an individual intercept (␤0i), the slope coefficient of the time score for time t for individual i (␤1i xti), and the residual symptoms of PTDS (eti) on time t for individual i. The equation term eti is computed by deviating the time-specific predictor (zti) from the regression line (␤1i xti) estimated separately (case by case) for each individual in the sample. The deviated measure, eti, is then the residual (i.e., the observed score minus expected value) from the regression of the timevarying predictor on time computed separately for each individual case, which then represents the variable for the within-person level of each predictor (i.e., Task, Goal, or Bond). The Level 2 between-person predictor represents variance due to interindividual differences in the time-varying predictor at the start of treatment, as shown in Equation A2: ϭ(␥00 ϩ u0i) ϩ (␥10 ϩ u1i)Ei(x ti). zbi ϭ ␤1i x 0i . In the Level 2 model, zbi is the between-person component of the time-varying predictor and is a function of individual differences in the time-varying predictor at the start of treatment (␤1i x0i). The equations for the model with main effects of the betweenperson and within-person predictors are presented in Equation A3: Level 1: ៮ ៮ eti ϭ (zti Ϫ zi) Ϫ (␥10 ϩ u1i)(x ti Ϫ x i) Level 2: ␤0i ϭ ␥00 ϩ u0i Equations for Proportion Reduction of Error at Each Level The proportion reduction of error for predicting the Level 1 outcome is (A2) R zti ϭ ␤0i ϩ ␤1i x ti ϩ eti In the Level 1 model, the individual-specific value of symptoms of PTSD (yti) is a function of an individual intercept (␤0i), the within-person effects of the time-varying predictor (␤1i xti), and the residual PTSD symptoms (eti) on time t for individual i. In the Level 2 model, the individual intercept (␤0i) is a function of a fixed intercept (␥00) and an individual-specific random intercept (u0i). The individual effects of slope (␤1i) is a function of the fixed effects in rate of change (␥10) and person-specific slope (u1i). (A3) R2 ϭ 1 – L1 ͩ residual variance more ϩ intercept variance more residual variance fewer ϩ intercept variance fewer ͪ . The proportion reduction of error for predicting the Level 2 outcome is R2 ϭ 1 – L2 ΂ residual variance more #Level 1 units residual variance fewer #Level 1 units ϩ intercept variance more ϩ intercept variance fewer ΃ . Received January 9, 2013 Revision received May 13, 2013 Accepted May 13, 2013 Ⅲ