Topics — What a work is about. Nucleation holds whatever reading the owner files here, whether or not an Anthology of the SOTA topic could also hold it; instruction for ML practice belongs to the anthology (ADR-048). The words are seeded from the clusters the reading-time triage of 2026-09-25 found, and grow by decision.
Psychometrics — measuring minds — scales, reliability and validity, latent-variable and network models of psychological constructs, and model-based estimation of cognitive parameters (ADR-025).
17 of 1352 THEORY documents. Back to the full index.
| # | Title | Summary | Status |
|---|---|---|---|
| THEORY-058 v2 | Mental disorders are maintained by self-reinforcing causal loops rather than expressing a latent common cause, which is how they can outlast their triggers | Three traditions that do not cite one another for it put self-sustaining loops where a latent cause would be. Borsboom (LIT-547) makes disorder an alternative stable state of a strongly connected symptom network, kept up by hysteresis after its trigger has gone. Salkovskis, Sighvatsson and Sigurdsson (LIT-538) make it a threat meaning protected by the behaviour it drives. Kendler (LIT-539) finds causal loops across levels, from genes to culture. The convergence is real; the evidence is indirect, simulated or clinical, and none of it is an individual time series showing hysteresis, so the account is Proposed. It says nothing about what starts a disorder. | Proposed |
| THEORY-061 | In evidence-accumulation tasks, goals and instructions act on a decision by setting the parameters of the accumulation (drift, starting point or baseline, and threshold), not by specifying the response, and this holds for when to act as well as for which | Bogacz et al. (LIT-569) define cognitive control, in a two-choice task, as the adjustment of drift, starting point and threshold to maximise a utility. Ratcliff and McKoon (LIT-576) find that the manipulations behave that way: speed or accuracy instructions are fitted by boundary separation alone, stimulus proportions mainly by starting point, difficulty by drift alone. Schurger et al. (LIT-605) carry the picture to the decision of when to move: the task goal raises the baseline toward threshold and leaves the moment to fluctuation. The convergence is real; the evidence is group fits on a few paradigms, so Proposed. It does not say who sets the parameters, or how. | Proposed |
| THEORY-074 | Implicit accessibility of the true self-concept is not what self-report scales of authenticity or of need satisfaction measure, though those two self-reports are not shown to be separate constructs | Schlegel, Hicks, Arndt & King (2009), LIT-606: true-self accessibility, the speed of “me” judgments for traits a person takes to be her true self, correlated .05 with AI-3 authenticity and .20 with basic need satisfaction (N = 124), and each predicted meaning in life independently. The authors read this as showing the constructs differ, and concede the difference is confounded with method, reaction time against self-report. The two self-reports correlated .52, and felt autonomy was only one third of the need-satisfaction composite, whose autonomy items include “I can pretty much be myself”. So the paper separates the implicit measure from both self-reports, and does not separate authenticity from autonomy. Proposed: one sample for the correlations, method unresolved. | Proposed |
| THEORY-475 | In simulations from the drift-diffusion model itself, with 12 subjects and 20 to 150 trials each, hierarchical Bayesian estimation recovers every DDM parameter with less error than per-subject maximum likelihood, χ²-quantile fitting or non-hierarchical Bayes, most of all at few trials; whether the gain survives misspecification, and how much of the power gain is the hierarchy’s, is not shown | Wiecki, Sofer & Frank (2013), LIT-563, read in NOTE-466. Experiment 1 gives the lowest trimmed recovery error for HB on every parameter, and HB beat ML on the same data sets in every condition. Experiments 2–3 report more power to detect drift differences and trial-wise covariate effects, but HB was judged by a posterior-interval test and the others by t-tests, and the clean comparison was omitted. All data were generated from DDMs, with sz = st = 0. It is the standard partial-pooling result shown for one model family, not a test of the DDM. | Active |
| THEORY-505 | Questionnaire items written for morality-as-cooperation’s seven domains form seven factors when rated for relevance to moral decisions, in the UK, the US and Turkey, but not when rated as agreement with moral statements, and the two formats cannot be pooled into seven domains | Curry, Jones Chesters & Van Lissa (2019), LIT-496, read in NOTE-412, and the independent Turkish validation by Yilmaz, Harma & Doğruyol (2021), LIT-513 (NOTE-397). Relevance items fit seven factors in all three samples (CFI .98 UK, .95 Turkey). Judgement items do not (CFI .87, .83), and a seven-domain model of both fails (CFI .73, .75). The items were written from the theory and selected for fit within each domain, several subscales correlate above .8, and no item is non-cooperative. This is evidence about the instrument’s structure, not about what morality contains. | Active |
| THEORY-613 | The Moral Foundations Questionnaire-2’s six factors, Care, Equality, Proportionality, Loyalty, Authority and Purity, fit across 19 nations, Equality and Proportionality are nearly uncorrelated in WEIRD nations, and the US split between individualizing and binding foundations appears in only three national networks | Atari, Haidt, Graham, Koleva, Stevens & Dehghani (2023), LIT-500, read in NOTE-398. 36 items, developed in 25 populations, fit six factors in 19 nations (ESEM CFI .979). Equality and Proportionality correlate .21 on average, from .47 to −.04. Purity alone fails intercept invariance, and it carries the headline that Purity is the least WEIRD foundation. The samples are online and about 200 per nation. | Proposed |
| THEORY-766 v2 | Brushing a seen rubber hand in synchrony with the hidden real hand, against asynchronous brushing, raises reports that touch is felt at the rubber hand and, more weakly and less reliably, that the rubber hand is one’s own; the proprioceptive drift toward the rubber hand that accompanies these reports is not a measure of them; the standard comparison conditions do not equalise participants’ expectations, so whether the stimulation or the expectation produces the report difference is open | Botvinick & Cohen (1998) and four replications read here give a synchrony effect on reports of referred touch and ownership. Rohde et al. (2011) and Abdulkarim & Ehrsson (2016) separate drift from ownership. The suggestion debate (Lush 2020; Lush et al. 2020; Ehrsson et al. 2022; Lush & Seth 2022; Roseboom & Lush 2022) shows that the standard controls do not equalise expectations. The ownership report is weaker than the referred-touch report and tracks trait suggestibility. Whether the synchronous stimulation, not expectation, produces the report difference is not settled. Promoted 2026-10-20 on the report differences measured (ADR-047 v3); the perceptual reading is THEORY-827. | Active |
| THEORY-892 v2 | Psychopathy as the PCL-R measures it is one dimensional construct rather than a type: its items form four correlated factors, interpersonal, affective, lifestyle and antisocial, explained by one superordinate factor, and its affective features are tied psychometrically, genetically and longitudinally to its antisocial ones, so that no one component is its essence | Hare and Neumann (2008), read in NOTE-1094. The PCL-R’s twenty items load on four correlated factors, which pair into Factor 1 (interpersonal, affective) and Factor 2 (lifestyle, antisocial) and are explained by one superordinate factor. The affective features are tied to the antisocial ones, so the authors deny that any component is the disorder’s core, and they read the evidence as dimensional. Filed for the record’s question on corporate psychopathy, which asks which of these features could be said of a corporation. | Proposed |
| THEORY-1148 | Single judges agree moderately on how pragmatically similar two spoken utterances are, mean pairwise correlation 0.45 to 0.72 by session, less when the utterances differ in words, so a mean over six to nine judges is much more reliable than any one judge | Ward and Marco (2024), LIT-1812, read in NOTE-1338 from the LREC-COLING 2024 PDF. Over 693 seed/re-enactment pairs in three sessions, mean pairwise Pearson agreement was 0.45, 0.72 and 0.66 (Table 2), and 0.80 in Session 2 when lexically different pairs were left out (p. 160). The Spearman–Brown reliability of the panel means, 0.88–0.96, is my computation. The judges talked during the session and knew each other, and all pairs are within one language, so the figure is neither a floor for a panel nor a value for translations. | Proposed |
| THEORY-1157 | Native-speaker crowd judges agree only modestly on the graded politeness of written requests in their own language, mean pairwise Spearman about 0.46 across nine languages, and the same bilinguals agree less on English requests, about 0.36 | Srinivasan and Choi (2022), LIT-1799, read in NOTE-1339 from the Findings of EMNLP 2022 PDF. Three Prolific annotators per item rated Wikipedia requests −3 to +3; mean pairwise Spearman within 20-item tasks was 0.46 on the 15 target-language items and 0.36 on the 5 English items (Table 2). The paper says own-language agreement is higher “across all nine languages”; Tamil (0.33 against 0.38) is the exception. The annotators had been screened for agreeing with an XLM-R classifier. | Proposed |
| THEORY-1311 | The BPNSFS frustration scales measure low need satisfaction in negatively keyed form, not need frustration in the sense of active thwarting, so the separation of frustration from low satisfaction in BPNSFS data, and frustration’s stronger link to ill-being there, is largely a keying artefact | Murphy et al. (2023), LIT-1978, read in NOTE-1530; Kermavnar et al. (2024), LIT-1988, read in NOTE-1509, adopt it. The item content and the keying structure (THEORY-1316) are shown. That the frustration scales are therefore low satisfaction is the authors’ conclusion, held “probably” and “likely primarily”, and by their own account cannot be decided with data in which keying and construct are confounded. Proposed for that step. | Proposed |
| THEORY-1316 | In eight samples pooled to 3,692 respondents, the covariance of the BPNSFS items is organized more by item keying than by need: factors sharing a keying direction correlate .64 to .75 while satisfaction and frustration of the same need correlate .17 to .38, a higher-order model grouping the six scales by keying fits better than one grouping them by need, and the frustration items carry almost no thwarting content | Murphy et al. (2023), LIT-1978, read in NOTE-1530 from the PMC author manuscript. Active as a statement of what their analyses show, checked against Tables 1–3 and the reported fits: the factor intercorrelations, the two higher-order models and their indices, the external correlates sorted by keying, and the items. It does not say that the frustration scales measure nothing but keying; the authors hold that this cannot be settled with existing data, and the inference is THEORY-1311’s. | Active |
| THEORY-1318 | Among 31 original self-report scales of psychological need fulfillment found by a 2023 systematic search, 21 measure satisfaction only, one measures frustration only and one thwarting only; all are self-reports of experience or of the perceived environment, the field uses satisfaction, frustration, support and thwarting interchangeably in places, and several frustration scales are negatively worded satisfaction items | Kermavnar, Avsec, Huang and Desmet (2024), LIT-1988, read in NOTE-1509 from the version of record. Active as a statement of what the review counted and found, checked against its §3 and Tables 2–4, within the scope it set (English, healthy general populations, title terms). It does not rank the scales, test their validity, offer a criterion for need-hood, or take a side on the two-dimensional model. | Active |
| THEORY-1327 | As Ryan and colleagues’ 2022 meta-review maps it, the meta-analytic evidence for self-determination theory is concentrated in cognitive evaluation, organismic integration and basic needs theory, is mostly cross-sectional and heterogeneous, and has gaps: no meta-analysis of relationships motivation theory or of need frustration, one or two of causality orientations and goal contents, and none that separates integrated from identified regulation | Ryan, Duineveld, Di Domenico, Ryan, Steward and Bradshaw (2022), LIT-1967, read in NOTE-1500 from the accepted manuscript on OSF. Active as a statement of what the review reports about the 60 meta-analyses it found to July 2021, which is checkable against its Tables 1 and 2 and its own summaries. It does not say the meta-analyses are sound, since the review grades none of them, and it is written by the theory’s co-founder, with several included meta-analyses by its authors. | Active |
| THEORY-1331 | Deci and Ryan’s General Causality Orientations Scale measures autonomy, control and impersonal orientations as three separately scored and largely independent dispositions, reliable and stable over two months, whose validity rests on correlations with self-report measures in small samples, with behaviour its weakest support | Deci and Ryan (1985), LIT-1955, read in NOTE-1531 from the scan on the SDT site. Active as a statement of what the scale is and what its founding paper showed: α .69–.74 and near-independent subscales in 636 students, retest .71–.78 in 51, and a nomological table in samples of 42 to 87. It does not say that orientations cause behaviour, or that the orientations parallel the three kinds of event of cognitive evaluation theory, which the paper asserts and its data only suggest. | Active |
| THEORY-1338 | In the Basic Psychological Need Satisfaction and Frustration Scale as Chen et al. built it, the distinction between satisfaction and frustration coincides with item direction: every satisfaction item states a positive experience and every frustration item an aversive one, so the better fit of six factors over three, which is the scale’s evidence that frustration is a separate dimension, is also what a keying or valence factor would produce | Chen et al. (2015), LIT-2000, read in NOTE-1527 from the publisher’s typeset copy in the ULB repository. Active as a statement about the instrument, checked against its Table 3 item by item and its model comparison (ΔSBSχ²(12) = 450.79, p. 226). It does not say the frustration dimension is an artefact, only that this evidence cannot tell it from one; that test is Murphy et al.’s (LIT-1978), unread. | Active |
| THEORY-1347 | In Patrick, Knee, Canevello and Lonsbary’s three correlational studies, perceiving that a romantic partner fulfils one’s needs for autonomy, competence and relatedness goes with better relationship quality, with relatedness the strongest unique predictor and the partner’s fulfilment adding to one’s own; but the measure does not separate the needs, its autonomy and competence subscales correlating .96 once corrected for unreliability | Patrick et al. (2007), LIT-1971, read in NOTE-1515 from the article on the SDT site. Active as a careful reading of reported results with their sizes and samples: eight pooled samples (N = 1,918), 66 couples, and 120 diarists with 908 disagreements, all self-report, mostly women students. It does not say need fulfilment causes relationship quality, that autonomous motivation mediates it (the mediation is cross-sectional), or that the three needs are distinct dimensions. | Active |