1. (1) To recognise and understand the development of PTSD following childbirth and baby loss.

2. (2) To understand how Ehlers and Clark’s (2000) cognitive model of PTSD can be applied to post-partum PTSD.

3. (3) To be able to apply cognitive therapy for PTSD to patients with perinatal PTSD, including traumatic baby loss through miscarriage or birth.

4. (4) To discover common personal meanings associated with birth trauma and baby loss and the steps to update them.

1. (1) To be able to apply Ehlers and Clark’s (2000) cognitive model to PTSD arising from bereavement trauma.

2. (2) To recognise how the core treatment components differ for PTSD associated with traumatic bereavement than for PTSD linked to trauma where there is no loss of life.

3. (3) To discover how to conduct imagery transformation for the memory updating procedure in CT-PTSD for loss trauma.

After reading this article, it is hoped that readers will be able to:

1. (1) Conceptualise why surveys can be useful in cognitive behavioural therapy.

2. (2) Implement collaborative and individualised survey design, delivery and feedback as part of a CBT intervention.

1. (1) To recognise common misconceptions about trauma-focused CBT for PTSD and the evidence against them.

2. (2) To widen understanding of the application of cognitive therapy for PTSD (CT-PTSD) to a broad range of presentations.

3. (3) To increase confidence in the formulation-driven, flexible, active and creative delivery of CT-PTSD.

Cognitive therapy for social anxiety disorder (CT-SAD) is recommended by NICE (2013) as a first-line intervention. Take up in routine services is limited by the need for up to 14 ninety-min face-to-face sessions, some of which are out of the office. An internet-based version of the treatment (iCT-SAD) with remote therapist support may achieve similar outcomes with less therapist time.

102 patients with social anxiety disorder were randomised to iCT-SAD, CT-SAD, or waitlist (WAIT) control, each for 14 weeks. WAIT patients were randomised to the treatments after wait. Assessments were at pre-treatment/wait, midtreatment/wait, posttreatment/wait, and follow-ups 3 & 12 months after treatment. The pre-registered (ISRCTN 95 458 747) primary outcome was the social anxiety disorder composite, which combines 6 independent assessor and patient self-report scales of social anxiety. Secondary outcomes included disability, general anxiety, depression and a behaviour test.

CT-SAD and iCT-SAD were both superior to WAIT on all measures. iCT-SAD did not differ from CT-SAD on the primary outcome at post-treatment or follow-up. Total therapist time in iCT-SAD was 6.45 h. CT-SAD required 15.8 h for the same reduction in social anxiety. Mediation analysis indicated that change in process variables specified in cognitive models accounted for 60% of the improvements associated with either treatment. Unlike the primary outcome, there was a significant but small difference in favour of CT-SAD on the behaviour test.

When compared to conventional face-to-face therapy, iCT-SAD can more than double the amount of symptom change associated with each therapist hour.

Psychological models of posttraumatic stress disorder (PTSD) and prolonged grief disorder (PGD) make predictions about the role of unhelpful coping strategies in maintaining difficulties by blocking self-correction of negative appraisals and memory integration following stressful life events like bereavement. However, few studies have tested these predictions directly.

We used counterfactually based causal mediation to assess whether unhelpful coping strategies mediated the relationship between (1) loss-related memory characteristics and/or (2) negative grief-related appraisals and symptoms of PGD, PTSD and depression using a three-wave longitudinal sample (N = 275). Appraisals and memory characteristics were measured at time point 1, unhelpful coping strategies at T2, and symptom variables at T3. Additionally, multiple mediation analyses within a structural equation modelling (SEM) framework assessed which types of coping strategies differentially mediated symptoms of PGD, PTSD and depression.

Coping strategies mediated the relationship between negative appraisals and memory characteristics and symptoms of PGD, PTSD and depression after adjusting for demographics and loss characteristics. Sensitivity analyses suggested that these results were most robust for PGD, followed by PTSD and then depression. Multiple mediation analyses suggested that all four subscales (avoidance, proximity seeking, loss rumination and injustice rumination) individually mediated the effect of memory characteristics and appraisals on PGD.

These results suggest that core predictions of the cognitive model for PTSD and the cognitive behavioural model of PGD are useful in predicting symptoms of post-loss mental health problems in the first 12–18 months after loss. Targeting unhelpful coping strategies is likely to reduce symptoms of PGD, PTSD and depression.

Cognitive therapies are developed on the principle that specific cognitive appraisals are key determinants in the development and maintenance of mental health disorders. It is likely that particular appraisals of the coronavirus pandemic will have explanatory power for subsequent mental health outcomes in the general public. To enable testing of this hypothesis we developed a questionnaire assessing coronavirus-related cognitions.

12 285 participants completed online a 46-item pool of cognitions about coronavirus and six measures of different mental health problems. The sample was randomly split into derivation and validation samples. Exploratory factor analyses determined the factor structure, selection of items, and model fit in the derivation sample. Confirmatory factor analysis (CFA) then tested this model in the validation sample. Associations of the questionnaire with mental health outcomes were examined.

The 26-item, seven-factor, Oxford Psychological Investigation of Coronavirus Questionnaire [TOPIC-Q] was developed. CFA demonstrated a good model fit (χ2 = 2108.43, df = 278, p < 0.001, comparative fit index (CFI) = 0.950, Tucker−Lewis index (TLI) = 0.942, root mean square error of approximation (RMSEA) = 0.033, standardized root mean square residual (SRMR) = 0.038). The factors were: cognitions about (1) safety and vulnerability, (2) negative long-term impact, (3) having the virus, (4) spreading the virus, (5) social judgment, (6) negative self, and (7) being targeted. The questionnaire explained significant variance in depression (45.8%), social anxiety (37.3%), agoraphobia (23.2%), paranoia (27.3%), post-traumatic stress disorder (57.1%), and panic disorder (31.4%). Cognitions about negative long-term impact had the greatest explanatory power across disorders.

TOPIC-Q provides a method to assess appraisals of the pandemic, which is likely to prove helpful both in longitudinal studies assessing mental health outcomes and in delivery of psychological therapy.

1. (1) To recognise moral injury where it arises alongside PTSD.

2. (2) To understand how Ehlers and Clark’s cognitive model of PTSD can be applied to moral injury.

3. (3) To be able to apply cognitive therapy for PTSD to patients with moral injury-related PTSD.

Catastrophic cognitive appraisals, similar to those in anxiety disorders, are implicated in depersonalisation, a form of dissociation. No scales exist to measure appraisals of dissociative experiences. Dissociation is common in psychosis. Misinterpretations of dissociative experiences may maintain psychotic symptoms. Therefore, assessing appraisals in this context may be valuable.

The primary aim was to develop a measure of key appraisals of dissociation in psychosis. Secondary aims were to test the relationship between appraisals and psychotic experiences (paranoia and hallucinations), and determine whether appraisals explain additional variance in psychotic symptoms above dissociative symptoms.

Fifty items were generated from transcripts of interviews with patients. The measure was developed and psychometrically validated via factor analysis of data from 9902 general population participants and 1026 patients with psychosis. Convergent validity, test–re-test reliability, and internal reliability were assessed. Regression analyses tested relationships with psychotic symptoms.

A 13-item single-factor measure was developed. Factor analysis indicated good model fit [χ2(65) = 247.173, comparative fit index (CFI) = 0.960, root mean square error of approximation (RMSEA) = 0.052]. The scale had good convergent validity with a rumination (non-clinical: r = 0.71; clinical: r = 0.73) and dissociation measure (r = 0.81; r = 0.80), high internal consistency (α = 0.93; α = 0.93), and excellent 1-week test–re-test reliability [intraclass correlation (ICC) = 0.90]. It explained variance in psychotic symptoms (paranoia: 36.4%; hallucinations: 35.0%), including additional variance compared with dissociation alone (paranoia: 5.3%; hallucinations: 2.3%).

The Cognitive Appraisals of Dissociation in Psychosis (CAD-P) measure is a psychometrically robust scale identifying appraisals of dissociative experiences in psychosis and is associated with the presence of psychotic experiences. It is likely to prove useful for clinical assessment and research.

Evidence-based treatment for panic disorder consists of disorder-specific cognitive behavioural therapy (CBT) protocols. However, most measures of CBT competence are generic and there is a clear need for disorder-specific assessment measures.

To fill this gap, we evaluated the psychometric properties of the Cognitive Therapy Competence Scale (CTCP) for panic disorder.

CBT trainees (n = 60) submitted audio recordings of CBT for panic disorder that were scored on a generic competence measure, the Cognitive Therapy Scale – Revised (CTS-R), and the CTCP by markers with experience in CBT practice and evaluation. Trainees also provided pre- to post-treatment clinical outcomes on disorder-specific patient report measures for cases corresponding to their therapy recordings.

The CTCP exhibited strong internal consistency (α = .79–.91) and inter-rater reliability (ICC = .70–.88). The measure demonstrated convergent validity with the CTS-R (r = .40–.54), although investigation into competence classification indicated that the CTCP may be more sensitive at detecting competence for panic disorder-specific CBT skills. Notably, the CTCP demonstrated the first indication of a relationship between therapist competence and clinical outcome for panic disorder (r = .29–.35); no relationship was found for the CTS-R.

These findings provide initial support for the reliability and validity of the CTCP for assessing therapist competence in CBT for panic disorder and support the use of anxiety disorder-specific competence measures. Further investigation into the psychometric properties of the measure in other therapist cohorts and its relationship with clinical outcomes is recommended.

1. (1) To learn how to deliver all of the core interventions of CT-SAD remotely.

2. (2) To learn novel ways of carrying out behavioural experiments remotely when some in-person social situations might not be possible.

1. (1) To recognise PTSD following admissions to intensive care units (ICUs).

2. (2) To understand how the ICU experience can lead to PTSD development.

3. (3) To understand how Ehlers and Clark’s (2000) cognitive model of PTSD can be applied to post-ICU PTSD.

4. (4) To be able to apply cognitive therapy for PTSD to patients with post-ICU PTSD.

Individual differences in cognitive responses to trauma may represent modifiable risk factors that could allow early identification, targeted early treatment and possibly prevention of chronic posttraumatic stress disorder (PTSD). Ehlers and Clark's cognitive model of PTSD suggests that negative appraisals, disjointed trauma memories, and unhelpful coping strategies maintain PTSD. These are thought to be influenced by cognitive processing during trauma. The aim of this study was to test this model prospectively with path analyses.

Participants (N = 828) were recruited from an emergency department following injury in a violent assault or road traffic collision and 700 participated in the 6-month assessments. Cognitive processing was assessed shortly after the event, negative appraisals, disjointed memories, and unhelpful coping strategies at 1 month, persistent PTSD symptom severity at 6 months, and early PTSD symptom severity at 2 weeks.

Cognitive variables, with trauma type and gender, explained 52% of the variance in PTSD symptom severity at 6 months. Including early symptom severity in the model did not explain more variance (53%). Early PTSD symptom severity, with trauma type and gender, only predicted 40%. Negative appraisals and disjointed memories predicted persistent symptom severity both directly and indirectly via unhelpful strategies. Peritraumatic processing predicted persistent symptom severity mainly indirectly. The effects of trauma type and gender were fully mediated by the cognitive factors.

The results are consistent with theoretically derived predictions and support cognitive factors as indicators of risk for chronic PTSD and as a target for the treatment and prevention of PTSD.

The cognitive process of worry, which keeps negative thoughts in mind and elaborates the content, contributes to the occurrence of many mental health disorders. Our principal aim was to develop a straightforward measure of general problematic worry suitable for research and clinical treatment. Our secondary aim was to develop a measure of problematic worry specifically concerning paranoid fears.

An item pool concerning worry in the past month was evaluated in 250 non-clinical individuals and 50 patients with psychosis in a worry treatment trial. Exploratory factor analysis and item response theory (IRT) informed the selection of scale items. IRT analyses were repeated with the scales administered to 273 non-clinical individuals, 79 patients with psychosis and 93 patients with social anxiety disorder. Other clinical measures were administered to assess concurrent validity. Test-retest reliability was assessed with 75 participants. Sensitivity to change was assessed with 43 patients with psychosis.

A 10-item general worry scale (Dunn Worry Questionnaire; DWQ) and a five-item paranoia worry scale (Paranoia Worries Questionnaire; PWQ) were developed. All items were highly discriminative (DWQ a = 1.98–5.03; PWQ a = 4.10–10.7), indicating small increases in latent worry lead to a high probability of item endorsement. The DWQ was highly informative across a wide range of the worry distribution, whilst the PWQ had greatest precision at clinical levels of paranoia worry. The scales demonstrated excellent internal reliability, test-retest reliability, concurrent validity and sensitivity to change.

The new measures of general problematic worry and worry about paranoid fears have excellent psychometric properties.