Measuring dissociation across adolescence and adulthood: developing the short-form Cernis Felt Sense of Anomaly scale (CEFSA-14).

BACKGROUND: Dissociation may be important across many mental health disorders, but has been variously conceptualised and measured. We introduced a conceptualisation of a common type of dissociative experience, 'felt sense of anomaly' (FSA), and developed a corresponding measure, the Cernis Felt Sense of Anomaly (CEFSA) scale. AIMS: We aimed to develop a short-form version of the CEFSA that is valid for adolescent and adult respondents. METHOD: Data were collected from 1031 adult NHS patients with psychosis and 932 adult and 1233 adolescent non-clinical online survey respondents. Local structural equation modelling (LSEM) was used to establish measurement invariance of items across the age range. Ant colony optimisation (ACO) was used to produce a 14-item short-form measure. Finally, the expected test score function derived from item response theory modelling guided the establishment of interpretive scoring ranges. RESULTS: LSEM indicated 25 items of the original 35-item CEFSA were age invariant. They were also invariant across gender and clinical status. ACO of these items produced a 14-item short-form (CEFSA-14) with excellent psychometric properties (CFI=0.992; TLI=0.987; RMSEA=0.034; SRMR=0.017; Cronbach's alpha=0.92). Score ranges were established based on the expected test scores at approximately 0.7, 1.25 and 2.0 theta (equivalent to standard deviations above the mean). Scores of 29 and above may indicate elevated levels of FSA-dissociation. CONCLUSIONS: The CEFSA-14 is a psychometrically valid measure of FSA-dissociation for adolescents and adults. It can be used with clinical and non-clinical respondents. It could be used by clinicians as an initial tool to explore dissociation with their clients.

The Oxford Positive Self Scale: psychometric development of an assessment of cognitions associated with psychological well-being.

BACKGROUND: Developing, elaborating, and consolidating positive views of the self is a plausible route to increased psychological well-being. We set out to provide an assessment of positive self-beliefs that could be used in research and clinical practice. METHODS: A non-probability online survey was conducted with 2500 UK adults, quota sampled to match the population for age, gender, ethnicity, income, and region. Exploratory factor analysis of a 94-item pool - generated with guidance from people with lived experience of mental health difficulties - was conducted to develop the Oxford Positive Self Scale (OxPos). The item pool was further reduced using regularised structural equation modelling (SEM) before confirmatory factor analysis. Optimal cut-off scores were developed using receiver operating characteristic curves. Additional validations were carried out with two further general population cohorts (n = 1399; n = 1693). RESULTS: A 24-item scale was developed with an excellent model fit [robust χ2 = 995.676; df = 246; CFI = 0.956; TLI = 0.951; RMSEA = 0.049 (0.047, 0.052); SRMR = 0.031]. The scale comprises four factors: mastery; strength; enjoyment; and character. SEM indicated that the scale explains 68.6% of variance in psychological well-being. The OxPos score was negatively correlated with depression (r = -0.49), anxious avoidance (r = -0.34), paranoia (r = -0.23), hallucinations (r = -0.20), and negative self-beliefs (r = -0.50), and positively correlated with psychological well-being (r = 0.79), self-esteem (r = 0.67), and positive social comparison (r = 0.72). Internal reliability and test-retest reliability were excellent. Cut-offs by age and gender were generated. A short-form was developed, explaining 96% of the full-scale variance. CONCLUSIONS: The new open access scale provides a psychometrically robust assessment of positive cognitions that are strongly connected to psychological well-being.

Listening to and Believing Derogatory and Threatening Voices.

BACKGROUND AND HYPOTHESIS: A plausible cause of distress for voice hearers is listening to and believing the threats and criticisms heard. Qualitative research indicates that patients have understandable reasons to listen. This study aimed to develop the understanding of distress using this listening and believing framework. Measures were developed of listening and believing voices and the reasons, and associations with distress tested. STUDY DESIGN: A cross-sectional study of patients hearing derogatory and threatening voices (N = 591). Listening and Believing-Assessment and Listening and Believing-Reasons item pools were completed, and assessments of distress. Exploratory and confirmatory factor analyses and structural equation modeling (SEM) were conducted. STUDY RESULTS: 52% (n = 307) of participants believed their voices most or all the time. Listening and believing had 4 factors: active listening, passive listening, believing, and disregarding. Higher levels of believing, active listening, and particularly passive listening were associated with higher levels of anxiety, depression, and voice distress. Reasons for listening and believing formed 7 factors: to better understand the threat; being too worn down to resist; to learn something insightful; being alone with time to listen; voices trying to capture attention; voices sounding like real people; and voices sounding like known people. Each type of reason was associated with active listening, passive listening, and believing. SEM showed that feeling worn down in particular accounted for listening and believing. Test-retest reliability of measures was excellent. CONCLUSIONS: A framework of listening and believing negative voices has the potential to inform the understanding and treatment of voice distress.

Developing a sentence level fairness metric using word embeddings.

Fairness is a principal social value that is observable in civilisations around the world. Yet, a fairness metric for digital texts that describe even a simple social interaction, e.g., 'The boy hurt the girl' has not been developed. We address this by employing word embeddings that use factors found in a new social psychology literature review on the topic. We use these factors to build fairness vectors. These vectors are used as sentence level measures, whereby each dimension reflects a fairness component. The approach is employed to approximate human perceptions of fairness. The method leverages a pro-social bias within word embeddings, for which we obtain an F1 = 79.8 on a list of sentences using the Universal Sentence Encoder (USE). A second approach, using principal component analysis (PCA) and machine learning (ML), produces an F1 = 86.2. Repeating these tests using Sentence Bidirectional Encoder Representations from Transformers (SBERT) produces an F1 = 96.9 and F1 = 100 respectively. Improvements using subspace representations are further suggested. By proposing a first-principles approach, the paper contributes to the analysis of digital texts along an ethical dimension.

Understanding agoraphobic avoidance: the development of the Oxford Cognitions and Defences Questionnaire (O-CDQ).

BACKGROUND: Many patients with mental health disorders become increasingly isolated at home due to anxiety about going outside. A cognitive perspective on this difficulty is that threat cognitions lead to the safety-seeking behavioural response of agoraphobic avoidance. AIMS: We sought to develop a brief questionnaire, suitable for research and clinical practice, to assess a wide range of cognitions likely to lead to agoraphobic avoidance. We also included two additional subscales assessing two types of safety-seeking defensive responses: anxious avoidance and within-situation safety behaviours. METHOD: 198 patients with psychosis and agoraphobic avoidance and 1947 non-clinical individuals completed the item pool and measures of agoraphobic avoidance, generalised anxiety, social anxiety, depression and paranoia. Factor analyses were used to derive the Oxford Cognitions and Defences Questionnaire (O-CDQ). RESULTS: The O-CDQ consists of three subscales: threat cognitions (14 items), anxious avoidance (11 items), and within-situation safety behaviours (8 items). Separate confirmatory factor analyses demonstrated a good model fit for all subscales. The cognitions subscale was significantly associated with agoraphobic avoidance (r = .672, p < .001), social anxiety (r = .617, p < .001), generalized anxiety (r = .746, p < .001), depression (r = .619, p < .001) and paranoia (r = .655, p < .001). Additionally, both the O-CDQ avoidance (r = .867, p < .001) and within-situation safety behaviours (r = .757, p < .001) subscales were highly correlated with agoraphobic avoidance. The O-CDQ demonstrated excellent internal consistency (cognitions Cronbach's alpha = .93, avoidance Cronbach's alpha = .94, within-situation Cronbach's alpha = .93) and test-re-test reliability (cognitions ICC = 0.88, avoidance ICC = 0.92, within-situation ICC = 0.89). CONCLUSIONS: The O-CDQ, consisting of three separate scales, has excellent psychometric properties and may prove a helpful tool for understanding agoraphobic avoidance across mental health disorders.

COVID-19 vaccine hesitancy in the UK: the Oxford coronavirus explanations, attitudes, and narratives survey (Oceans) II.

BACKGROUND: Our aim was to estimate provisional willingness to receive a coronavirus 2019 (COVID-19) vaccine, identify predictive socio-demographic factors, and, principally, determine potential causes in order to guide information provision. METHODS: A non-probability online survey was conducted (24th September-17th October 2020) with 5,114 UK adults, quota sampled to match the population for age, gender, ethnicity, income, and region. The Oxford COVID-19 vaccine hesitancy scale assessed intent to take an approved vaccine. Structural equation modelling estimated explanatory factor relationships. RESULTS: 71.7% (n=3,667) were willing to be vaccinated, 16.6% (n=849) were very unsure, and 11.7% (n=598) were strongly hesitant. An excellent model fit (RMSEA=0.05/CFI=0.97/TLI=0.97), explaining 86% of variance in hesitancy, was provided by beliefs about the collective importance, efficacy, side-effects, and speed of development of a COVID-19 vaccine. A second model, with reasonable fit (RMSEA=0.03/CFI=0.93/TLI=0.92), explaining 32% of variance, highlighted two higher-order explanatory factors: 'excessive mistrust' (r=0.51), including conspiracy beliefs, negative views of doctors, and need for chaos, and 'positive healthcare experiences' (r=-0.48), including supportive doctor interactions and good NHS care. Hesitancy was associated with younger age, female gender, lower income, and ethnicity, but socio-demographic information explained little variance (9.8%). Hesitancy was associated with lower adherence to social distancing guidelines. CONCLUSIONS: COVID-19 vaccine hesitancy is relatively evenly spread across the population. Willingness to take a vaccine is closely bound to recognition of the collective importance. Vaccine public information that highlights prosocial benefits may be especially effective. Factors such as conspiracy beliefs that foster mistrust and erode social cohesion will lower vaccine up-take.

The revised Green et al., Paranoid Thoughts Scale (R-GPTS): psychometric properties, severity ranges, and clinical cut-offs.

BACKGROUND: The Green et al., Paranoid Thoughts Scale (GPTS) - comprising two 16-item scales assessing ideas of reference (Part A) and ideas of persecution (Part B) - was developed over a decade ago. Our aim was to conduct the first large-scale psychometric evaluation. METHODS: In total, 10 551 individuals provided GPTS data. Four hundred and twenty-two patients with psychosis and 805 non-clinical individuals completed GPTS Parts A and B. An additional 1743 patients with psychosis and 7581 non-clinical individuals completed GPTS Part B. Factor analysis, item response theory, and receiver operating characteristic analyses were conducted. RESULTS: The original two-factor structure of the GPTS had an inadequate model fit: Part A did not form a unidimensional scale and multiple items were locally dependant. A Revised-GPTS (R-GPTS) was formed, comprising eight-item ideas of reference and 10-item ideas of persecution subscales, which had an excellent model fit. All items in the new Reference (a = 2.09-3.67) and Persecution (a = 2.37-4.38) scales were strongly discriminative of shifts in paranoia and had high reliability across the spectrum of severity (a > 0.90). The R-GPTS score ranges are: average (Reference: 0-9; Persecution: 0-4); elevated (Reference: 10-15; Persecution: 5-10); moderately severe (Reference: 16-20; Persecution:11-17); severe (Reference: 21-24; Persecution: 18-27); and very severe (Reference: 25+; Persecution: 28+). Recommended cut-offs on the persecution scale are 11 to discriminate clinical levels of persecutory ideation and 18 for a likely persecutory delusion. CONCLUSIONS: The psychometric evaluation indicated a need to improve the GPTS. The R-GPTS is a more precise measure, has excellent psychometric properties, and is recommended for future studies of paranoia.

The assessment of paranoia in young people: Item and test properties of the Bird Checklist of Adolescent Paranoia.

BACKGROUND: Precise assessment tools for psychotic experiences in young people may help identify symptoms early and facilitate advances in treatment. In this study we provide an exemplar - with a paranoia scale for youth - for improving measurement precision for psychotic experiences using item response theory (IRT). We evaluate the psychometric properties of the new measure, test for measurement invariance, and assess its potential for computerised adaptive testing (CAT). METHOD: The 18-item Bird Checklist of Adolescent Paranoia (B-CAP) was completed by 1102 adolescents including 301 patients with mental health problems and 801 from the general population. After excluding outliers (n = 10), IRT was used to examine item properties, test reliability, and measurement invariance. The properties of an adaptive B-CAP were assessed using a simulation of 10,000 responses. RESULTS: All B-CAP items were highly discriminative (a = 1.14-2.77), whereby small shifts in paranoia led to a higher probability of item endorsement. Test reliability was high (a > 0.90) across a wide range of paranoia severity (θ = -0.45-3.36), with the greatest precision at elevated levels. All items were invariant for gender, age, and population groups. The simulated adaptive B-CAP performed with high accuracy and required only 5-6 items at higher levels of paranoia severity. CONCLUSIONS: The B-CAP is a reliable assessment tool with excellent psychometric properties to assess both non-clinical and clinical levels of paranoia in young people, with potential as an efficient adaptive test. In future, these approaches could be used to develop a multidimensional CAT to assess the full range of psychotic experiences in youth.

The Dunn Worry Questionnaire and the Paranoia Worries Questionnaire: new assessments of worry.

BACKGROUND: 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. METHODS: 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. RESULTS: 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. CONCLUSIONS: The new measures of general problematic worry and worry about paranoid fears have excellent psychometric properties.

Development of a measure of genome sequencing knowledge for young people: The kids-KOGS.

Genome sequencing (GS) is increasingly being used to diagnose rare diseases in paediatric patients; however, no measures exist to evaluate their knowledge of this technology. We aimed to develop a robust measure of knowledge of GS (the kids-KOGS') suitable for use in the paediatric setting as well as for general public education. The target age was 11 to 15 year olds. An iterative process involving six sequential stages was conducted to develop a set of draft true/false items. These were then administered to 539 target-age school pupils (mean 12.8; SD ± 1.3), from the United Kingdom. Item-response theory was used to confirm the psychometric suitability of the candidate items. None of the Items was identified as misfits. All 10 items performed well under the two-parameter logistic model. The internal consistency of the test was 0.84 (Cronbach alpha value) indicating excellent reliability. The mean kids-KOGS score in the sample overall was 4.24 (SD; 2.49), where 0 = low knowledge and 10 = high knowledge. Age was positively associated with score in a multivariate linear regression. The kids-KOGS is a short and reliable tool that can be used by researchers and healthcare professionals offering GS to paediatric patients. Further validation in a clinical setting is required.

Adolescent Paranoia: Prevalence, Structure, and Causal Mechanisms.

BACKGROUND: Adolescence can be a challenging time, characterized by self-consciousness, heightened regard for peer acceptance, and fear of rejection. Interpersonal concerns are amplified by unpredictable social interactions, both online and offline. This developmental and social context is potentially conducive to the emergence of paranoia. However, research on paranoia during adolescence is scarce. METHOD: Our aim was to examine the prevalence, structure, and probabilistic causal mechanisms of adolescent paranoia. A representative school cohort of 801 adolescents (11-15 y) completed measures of paranoia and a range of affective, cognitive, and social factors. A Bayesian approach with Directed Acyclic Graphs (DAGs) was used to assess the causal interactions with paranoia. RESULTS: Paranoid thoughts were very common, followed a continuous distribution, and were hierarchically structured. There was an overall paranoia factor, with sub-factors of social fears, physical threat fears, and conspiracy concerns. With all other variables controlled, DAG analysis identified paranoia had dependent relationships with negative affect, peer difficulties, bullying, and cognitive-affective responses to social media. The causal directions could not be fully determined, but it was more likely that negative affect contributed to paranoia and paranoia impacted peer relationships. Problematic social media use did not causally influence paranoia. CONCLUSIONS: There is a continuum of paranoia in adolescence and occasional suspicions are common at this age. Anxiety and depression are closely connected with paranoia and may causally contribute to its development. Paranoia may negatively impact adolescent peer relationships. The clinical significance of paranoia in adolescents accessing mental health services must now be established.