ORIGINAL RESEARCH published: 18 November 2021 doi: 10.3389/fpsyt.2021.659835 Decentering, Acceptance, and Non-Attachment: Challenging the Question “Is It Me?” Joaquim Soler 1,2,3*, Jesus Montero-Marin 4 , Elisabet Domínguez-Clavé 1,3,5 , Sara González 1,3 , Juan Carlos Pascual 1,2,3 , Ausiàs Cebolla 6,7 , Marcelo Demarzo 8 , Bhikkhu Analayo 9,10 and Javier García-Campayo 11,12 1 Servei de Psiquiatria, Hospital de la Santa Creu i Sant Pau, Barcelona, Spain, 2 Department of Psychiatry and Forensic Medicine, Universitat Autònoma de Barcelona (UAB), Barcelona, Spain, 3 Centro de Investigación Biomédica en Red de Salud Mental, CIBERSAM, Madrid, Spain, 4 Department of Psychiatry, Warneford Hospital, University of Oxford, Oxford, United Kingdom, 5 Department of Pharmacology and Therapeutics, Universitat Autònoma de Barcelona (UAB), Barcelona, Spain, 6 Departamento de Personalidad, Evaluación y Tratamientos Psicológicos, University of Valencia UV, València, Spain, 7 CIBER Physiopathology of Obesity and Nutrition (CIBEROBN), Madrid, Spain, 8 Mente Aberta – Brazilian Center for Mindfulness and Health Promotion, Universidade Federal de São Paulo (UNIFESP), São Paulo, Brazil, 9 Barre Center of Buddhist Studies (BCBS), Barre, MA, United States, 10 Numata Center for Buddhist Studies, University of Hamburg, Hamburg, Germany, 11 Red de Investigación de Actividades Preventivas y Promoción de la Salud (RedIAPP), Zaragoza, Spain, 12 Miguel Servet Hospital, University of Zaragoza, Zaragoza, Spain Edited by: Rakesh Pandey, Banaras Hindu University, India Among mindfulness measures the three constructs acceptance, decentering, and Reviewed by: non-attachment are psychometrically closely related, despite their apparent semantic Durgesh Kumar Upadhyay, Mahatma Gandhi Kashi differences. These three facets present robust psychometric features and can be Vidyapith, India considered core themes in most “third wave” clinical models. The aim of the present study Darpan Kaur, Mahatma Gandhi Missions Medical was to explore the apparently different content domains (acceptance, decentering, and College and Hospital, India non-attachment) by administering various psychometric scales in a large sample of 608 Anubhuti Dubey, volunteers. Resilience and depression were also assessed. Exploratory and confirmatory Deen Dayal Upadhyay Gorakhpur University, India factor analyses performed in two randomly selected subsamples showed a bifactor *Correspondence: approximation. The explained common variance suggested a unidimensional nature for Joaquim Soler the general factor, with good psychometric properties, which we named “Delusion of jsolerri@santpau.cat Me” (DoM). This construct is also strongly correlated with resilience and depression, and Specialty section: appears to be a solid latent general construct closely related to the concept of “ego.” This article was submitted to DoM emerges as a potentially transdiagnostic construct with influence on well-being Psychological Therapies, a section of the journal and clinical indexes such as resilience and depression. Further studies should analyze Frontiers in Psychiatry the potential utility of this new construct at a therapeutic level. Received: 28 January 2021 Keywords: mindfulness, acceptance, decentering, non-attachment, depression, resilience Accepted: 11 October 2021 Published: 18 November 2021 Citation: INTRODUCTION Soler J, Montero-Marin J, Domínguez-Clavé E, González S, In recent years, mindfulness has become an important health-related concept, and the absence Pascual JC, Cebolla A, Demarzo M, of mindfulness has been consistently associated with clinical symptoms, both medical and Analayo B and García-Campayo J (2021) Decentering, Acceptance, and psychological (1, 2). By contrast, the presence of mindfulness has been positively correlated Non-Attachment: Challenging the with quality of life and well-being (3–6). Traditionally, mindfulness is considered to contain Question “Is It Me?”. two main components: (1) self-regulation of attention and (2) an attitude of acceptance (7). Front. Psychiatry 12:659835. Currently available measures of mindfulness are highly heterogeneous assessing as few as a single doi: 10.3389/fpsyt.2021.659835 characteristic of mindfulness (8) to up to nine distinctive features (9). Given this heterogeneity, Frontiers in Psychiatry | www.frontiersin.org 1 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment together with other controversial aspects such as diverse Acceptance and Action Questionnaire-II [AAQ-II, (30)] and the theoretical backgrounds, differences in the definitions used in FFMQ non-judging facet (27, 31). Feliu-Soler et al. (27) found a these scales, the trait or state debate, and the question of to what close association between non-attachment and decentering, and extent mindfulness is trainable (10–13), it is not surprising that Allen (32) suggested that decentering appears to be a quality relationship between the various mindfulness scales is generally embedded in non-attachment. Furthermore, non-attachment poor, particularly in the case of dispositional mindfulness and to the self—which refers to a decrease in self-referential non-experienced populations (14). Importantly, this lack of processing or Experiential Selfless Processing (33), that is, consistency between these instruments does not apply to all processing the present moment subjective experience without of the sub-facets and by-products of mindfulness practice. In self-referentiality—has been identified as a core mechanism in fact, certain aspects such as acceptance, decentering, and non- mindfulness training (34–36). attachment are all highly correlated. It could be argued that acceptance, decentering, and Several instruments are available to assess measures for each non-attachment are all intimately connected, not only of these three domains. In the most commonly used mindfulness psychometrically but also theoretically. “If one grasps onto the questionnaire—the “Five Facets Mindfulness Questionnaire” true existence object [by fusion], this gives rise to attachment” [FFMQ, (15)]—acceptance is best represented by accepting (37). The unaware identification with the content of perception without judgment, which refers to taking a non-evaluative and thought is one contemporary conceptualization of the stance toward the feelings and thoughts. Nevertheless, the term “ego” (38). This leads to the confusion of our mental developer of the FFMQ originally suggested that non-reacting representations with ourselves. As the self-verification theory together with non-judging of inner experience both represent suggests, people strongly identify with those representations and acceptance (16). Several studies have shown a significant overlap foster, protect, and defend those self-images (39). between the non-reacting and the decentering constructs, which When we do not accept something, we are judging it to consistently correlate with each other, even more so than with be good or bad, and this judgment is fused with a particular the non-judging facet (10, 17, 18). Likewise, the non-judging desired outcome that seeks our “own” benefit. Some core themes subscale strongly correlates with other measures of acceptance, in mindfulness-based programs (MBPs) partially represent this such as experiential avoidance in the “Acceptance and Action overlapping nature, which can be seen in the concepts of Questionnaire-II” (AAQ-II) and the acceptance subscale in the “aversion” and “craving” in mindfulness-based cognitive therapy “Philadelphia Mindfulness Scale” (PHLMS) (1, 19, 20). (MBCT), the idea of “letting go” in both mindfulness-based Decentering, also referred to as defusion, reification, or meta- stress reduction (MBSR) and MBCT, and in radical acceptance awareness in the literature, is commonly assessed with the principles and techniques such as “willingness vs. willfulness” in Experiences Questionnaire [EQ, (21)]. However, decentering dialectical behavior therapy (DBT). In all cases, one relinquishes is also evaluated in other instruments, including the Toronto involvement and frees oneself from the need for things to Mindfulness Scale [TMS, (22)], the Southampton Mindfulness be different (40–42). These attitudinal components, which are Questionnaire [SMQ, (23)], and the Comprehensive Inventory closely related to acceptance, could be more stable and reliable of Mindfulness Experience [CHIME, (20)]. Decentering has been characteristics than the attentional component of mindfulness, defined as “the ability to observe one’s thoughts and feelings in which appear to be mood dependent (2). Given the theoretical a detached manner, as temporary events in the mind, as neither and psychometric correlation among acceptance, decentering, necessarily true nor reflections of the self ” (24). Studies have and non-attachment, we hypothesized that these concepts may, shown that a low level of decentering can, by itself, predict an in fact, be different semantic expressions, or the building blocks, earlier relapse in individuals who have recently recovered from of the same latent construct. Therefore, we believe that this major depression (25). Some reports also suggest that decentering new construct could be central to several views of “ego” where may be impaired in individuals with borderline personality fusion with mental content is a key feature, such as the self-as- disorder, eating behavior disorder, or cocaine dependence (26). object (38), the analytical self (43), the narrative self (44, 45), Decentering is closely related to measures of acceptance such or the self-as-content (46). If this construct is reliable, it would as experiential avoidance and non-acceptance, both included in allow us to better understand a form of self that, when becomes the Difficulties in Emotion Regulation Scale (DERS), and with overidentified, can be the cause of suffering (38). The idea of non-judging from the FFMQ (27, 28). a unified self with a sense of agency relies on a coherent set A final but equally important element is non-attachment, of beliefs, intentions, and behaviors, and is also represented in which has its origins in Buddhism. Non-attachment can be several contemplative traditions (47) that have been taken as a defined as the relative absence of fixation on ideas, images, or source of inspiration for some evidence based contemporaneous sensory objects and as an absence of internal pressure to obtain, psychotherapies (40–42). Those traditions advocate for the idea hold, avoid, or change circumstances or experiences (29). This of non-self and actions performed, or motivated by lack of construct has been positively associated with meditation practice desire (48–50). In all three cases attachment, fusion, and non- (total weekly hours and frequency of practice), suggesting acceptance are part of the equation. that regular mindfulness practice may promote non-attachment To explore this hypothesis, we sought to analyze the possible (27). The Non-Attachment Scale [NAS, (29)] is an accurate unidimensionality of a general factor underlying acceptance, representation of non-attachment, and as with decentering, it decentering, and non-attachment using the non-judging facet is closely correlated with acceptance (29), as assessed with the included in the FFMQ, EQ, and NAS scales. Second, we Frontiers in Psychiatry | www.frontiersin.org 2 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment TABLE 1 | General socio-demographic variables of the participants according to subsample. Variables/samples Total sample Subsample 1 Subsample 2 p (n = 608) (n = 304) (n = 304) Range Mean SD Mean SD Mean SD Age, years* 17–75 41.32 11.13 40.91 11.52 41.72 10.72 0.379 Category Freq. % Freq. % Freq. % Gender** Female 374 63.2 191 64.5 183 61.8 0.495 Male 218 36.8 105 35.5 113 38.2 Education** Primary 11 1.9 6 2.0 5 1.7 9.982 Secondary 117 19.8 57 19.3 60 20.3 Diploma 86 14.5 45 15.2 41 13.9 Degree 300 50.7 150 50.7 150 50.7 Ph.D. 78 13.2 38 12.8 40 13.5 Meditation** Yes 335 56.6 160 54.1 175 59.1 0.214 No 257 43.4 136 45.9 121 40.9 SD, standard deviation; Freq, frequencies; %, percentages. *18 missing values (9 in subsamples 1 and 9 in subsample 2). **16 missing values (8 in subsamples 1 and 8 in subsample 2). also hypothesized that this new general construct, which may and time taken was around 20 min. A total of 850 subjects subsume these three facets, would be positively correlated voluntarily agreed to participate; of these, 608 (72%) completed with well-being and negatively correlated with psychopathology. all of the requirement instruments and questionnaires. Two Therefore, the aim of the present study was to define this randomly selected halves of this sample, with a size of n1 = 304 construct and then to perform a preliminary test of its and n2 = 304, supposing a null hypothesis that RMSEA would nomological validity compared with an important positive be equal to or <0.04 if the true value was 0.06 (close fit) and an psychology index, resilience, and to depressive symptoms. alpha equal to 0.05 level, under the more conservative model tested (i.e., only one factor that covers all the items), produces a power coefficient of 0.99 for the exploratory and confirmatory MATERIALS AND METHODS factor analyses. The socio-demographic characteristics of the A Cross-sectional study was developed with a convenience sample and the randomly-selected subsamples are described in sample of Spanish general population participants that were Table 1. included if: (1) aged >18 years, (2) able to understand Spanish, and (3) signed informed consent. Measures First, we collected information on age, gender (male, female), Participants and Procedure education (primary, secondary, Associate’s degree, Bachelor’s Participants were recruited through several Spanish websites. degree, Ph.D.), and previous experience with meditation (yes An internet-based commercial system specifically designed or no). to recruit survey samples was used to recruit our sample The Spanish version of the FFMQ (16, 19) is a tool designed (www.surveymonkey.com; Portland, OR, USA). A link to evaluate mindfulness. For the purposes of the present study, containing the assessment protocol was posted on mindfulness we used only the “non-judging of inner experience” dimension and meditation focused portals and several mindfulness of the FFMQ. This dimension comprises eight items (e.g., associations, sanghas, and Zen monasteries. The non-meditative “I make judgments about whether my thoughts are good or convenience samples were enrolled accessing to the link placed bad”) that are designed to assess an individual’s ability to in scientific and psychology research sites and non-professional take a non-evaluative stance toward experience. In view of the social networks (e.g., Facebook). The study protocol was existing debate surrounding the most appropriate measure of approved by the corresponding regional health authority, the acceptance—traditionally associated with non-judging and non- Aragon Ethics Committee (CEICA), Spain (registry: PI12/00083) reactivity FFMQ subscales (18, 20, 51)—we elected to use only and all participants signed a consent form indicating their the first subscale as this is more closely associated with other willingness to participate. Prospective participants were invited acceptance measures than the FFMQ non-reactivity subscale, to participate voluntarily in the study and were told that both in clinical and non-clinical samples (1). Participants were there would be no monetary compensation. Participants were asked to rate each sentence on a five-point scale ranging from informed about the purpose of the study and informed that 1 (“never or very rarely true”) to 5 (“very often or always all responses would be treated confidentially. Scales were filled true”), with higher scores indicating higher levels of non- online, with the possibility of using mobile phones or computers, judging of inner experience. The Spanish language version of the Frontiers in Psychiatry | www.frontiersin.org 3 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment FFMQ scale and subscales has shown appropriate psychometric greater depressive symptoms. The composite reliability of the properties (19). The composite reliability of non-judging inner depression subscale in the present study was adequate, with a experience in the present study sample was ω = 0.93. value of ω = 0.90. The EQ (21, 26) is used to assess the ability of an individual to The selection of 24 items, coming from FMMQ, EQ and NAS- observe their own thoughts and emotions as temporary objects 7, which are the components of the new construct studied, are of the mind, captured under the concept of “decentering.” The specified in the Supplementary Material. EQ contains 11 items designed to measure a metacognitive ability known as “decentering,” defined as the capacity to observe one’s Data Analyses thoughts and emotions in a detached manner, considering them Sociodemographic data were described as means with standard transient events of the mind (e.g., “I can separate myself from deviation (SD) and frequencies (percentages) according to the my thoughts and feelings”). Two original items from the EQ (“I statistical distribution of each variable. Possible differences can treat myself kindly” and “I can slow my thinking at times between two randomly selected subsamples were tested using of stress”) were not included in the present analysis because the t-tests for independent groups and χ 2 (or Fisher’s exact divergent meaning of these items appears to be more closely test when necessary) tests depending on the shape of the related to the self-compassion construct (52). Thus, in the present score distributions of measures. Individual item distributions study, we used a nine-item version of the EQ scale. All items on were described in each independent random subsample using the EQ are scored on a five-point scale, ranging from 1 (“never”) means with SD. Mardia’s statistics were calculated to evaluate to 5 (“all the time”), with higher scores indicating a greater the multivariate behavior of the items. We verified the decentering capacity. The Spanish language version of the EQ Kaiser–Meyer–Olkin (KMO) sampling adequacy values, the has shown adequate psychometric properties (26). The composite Bartlett’s test of sphericity on the redundancy levels, and the reliability of the final nine decentering items in the present study matrix determinants to rule out multi-colinearity problems. was ω = 0.87. To determine the underlying factorial structure of the items, The NAS-7 (31) is a 7-item measure extracted from the we used Schwartz’s Bayesian information criterion (BIC) as a original 30-item NAS. The original NAS-30 (29) was empirically dimensionality test in both subsamples. derived from a pool of items obtained from Buddhist texts about Item–rest correlations were calculated, and an exploratory non-attachment. The NAS-7 is a unidimensional short form that factor analysis (EFA) was performed in the first randomly measures the absence of fixation on ideas, images, or sensory selected subsample. The robust maximum likelihood (RML) objects, as well as the absence of internal pressure to obtain, method, with correction for robust mean and variance-scaled, hold, avoid, or change circumstances or experiences. Items (e.g., was used for factor extraction. We tested the following models: “I can let go of regrets and feelings of dissatisfaction about the (a) one first-order factor, as the simplest solution, taken as a past”) are scored on a six-point scale ranging from 1 (“strongly reference, (b) three correlated first-order factors, maintaining disagree”) to 6 (“strongly agree”), with higher scores indicating the origin of the items, (c) the Schmid–Leiman solution, as greater levels of non-attachment. NAS-7 has shown good internal an exploratory second order factor approximation, and (d) consistency for both meditators and non-meditators (27). The exploratory bifactor, allowing for the existence of a general composite reliability of non-attachment in the present study was common factor and three orthogonal sub-factors. Standardized ω = 0.85. factor loadings (λ), uniqueness terms (δ), latent inter-factor The Connor-Davidson Resilience Scale-10 [CD-RISC-10 (53, correlations (ϕ), and discrepancy values—as unstandardized 54)] is a self-administered 10-item measure designed to assess residual covariance estimates were also considered. Raw loading a broad construct of resilience (e.g., “I can deal with whatever matrices were rotated using the Promin procedure, which allows comes”), a protective factor against mental problems that is factors to be oblique to maximize factor simplicity, without positively associated with adaptive coping. Items from the CD- assuming that all the variables are pure measures of a single RISC-10 range from 0 (“not true at all”) to 4 (“true nearly all the dimension (57). We evaluated factorial simplicity by means of time”). There is a total scale score, with higher scores indicating (a) the index of factor simplicity (IFS), (b) the scale fit index greater resilience. The original study on the development of (SFI), (c) Bentler’s scale-free matrix measure, and (d) hyperplane the CD-RISC in the general population and inpatients provided counts. IFS and SFI values ≥0.80 are appropriate; Bentler’s support for the internal consistency, test–retest reliability, and measure ranges from 0 (very complex structures) to 1 (very validity of this scale (53). The composite reliability of the CD- simple ones). Hyperplane counts (loadings essentially zero except RISC-10 scale in the present study was good, with a value of ω for random error) were estimated through the −0.15/+0.15 = 0.91. interval and through the Kaiser and Cerny procedure (58). We The Depression Anxiety Stress Scale, short form [DASS-21 explored closeness to unidimensionality by means of the item (55, 56)], is a self-administered, 21-item instrument with three residual absolute loadings (MIREAL) and the explained common subscales, which assesses depressive symptomatology, anxiety, variance (ECV). MIREAL is a measure of departure from and stress. In the present study, we used only the depression unidimensionality, with values <0.30 indicating no substantial subscale (e.g., “life is meaningless”). The items in that subscale bias if a unidimensional solution is adopted (59, 60). ECV assess the severity/frequency of symptoms over the previous week represents the proportion of common variance attributable to on a four-point scale ranging from 0 (“nothing applicable to the general factor, which should fall within the 0.70–0.85 range me”) to 3 (“very applicable to me”), with higher scores indicating if a solution is to be accepted as unidimensional (61). The Frontiers in Psychiatry | www.frontiersin.org 4 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment effectiveness of factor scores was assessed by using the factor Finally, to assess the nomological validity of the proposed determinacy index (FDI). FDI is the correlation between the general factor, we evaluated the possible links between the general factor score estimates and the levels on the latent factors they factor and the constructs of resilience and depression. To do estimate (62), and values of around 80 are considered adequate so, we constructed a structural equation model (SEM) using the (63). Construct replicability is the proportion of the factor ML method. We calculated the raw inter-factor standardized variance that can be accounted for by its indicators, which was latent correlations between the general factor and the other measured by the H index (bounded between 0 and 1) with values constructs, as well as the adjusted standardized links between the considered appropriate when ≥0.80 (61). latent variables involved in the model. The explained variance of A confirmatory factor analysis (CFA) using the maximum the latent variables (R2 ), as well as the unstandardized residual likelihood (ML) method on the second randomly selected covariance, was estimated. subsample was conducted to cross-validate the factor structures We conducted analyses for participants with non-missing obtained in the EFA.The goodness-of-fit was assessed by data. The tests used were bilateral, and the significance chi-square (χ 2 ), chi-square/degrees of freedom (χ 2 /df), the level was α <0.05. The Factor (v.10.9.02) and AMOS (v.20) comparative fit index (CFI), the Tucker–Lewis index (TLI), the software packages for Windows were used to perform the root mean square error of approximation (RMSEA), and the statistical analysis. standardized root mean square residual (SRMR). Chi-square is highly sensitive to sample size, so χ 2 /df was used, which RESULTS indicates a good fit when <5 and excellent fit <3 (64, 65). CFI analyzes the model fit by examining the discrepancy between Sociodemographic Data the data and the hypothesized model adjusted for the sample Table 1 shows the sociodemographic characteristics of the study size, and TLI analyzes the discrepancy between the χ 2 value of participants. No significant differences were observed between the hypothesized model and the χ 2 value of the null model, the two randomly selected subsamples in terms of age, gender, penalizing for the number of parameters. Both CFI and TLI educational level, or meditation experience. incremental measures indicate adequate fit when the value is >0.90, and an excellent fit when >0.95 (64, 66). RMSEA is a Exploratory Factor Analyses measurement of the error of approximation to the population, Descriptive statistics for all items (subsample 1) are shown in and SRMR is the standardized difference between the observed Table 2. The item-rest correlations were in the same direction and the predicted covariance. Both RMSEA and SRMR absolute for all items, with values ranging from 0.42 (item NA5) to 0.67 measures indicate adequate fit with values <0.08 and excellent fit (item NJ3). The results of the BIC dimensionality test suggested a when <0.06 (64, 66). one-factor solution (Table 3), which explained 39.3% of the total The omega composite reliability for the total scale (ω), as variance, with loadings (Table 2) between 0.41 (item NA5) and well as for each subscale (ωS ), were calculated, which may be 0.74 (items NJ2 and NJ5). The ECV was 0.80, and the MIREAL interpreted as the square of the correlation between the scale was 0.29; thus, the unidimensionality of data was supported in (ω)—or subscale (ωS )—score, and the latent variable common the first subsample. Factor determinacy (FDI = 0.97), omega to the corresponding indicators (67). We also estimated the reliability (ω = 0.93), and construct replicability (H = 0.94) were omega hierarchical (ωH ), as the proportion of reliable variance all very good, but the goodness of model-data fit for the one- in total scores that can be attributed to the single general factor solution was not adequate in terms of RMSEA and SRMR factor, as well as the omega hierarchical subscale (ωHS), as values (Table 4). the proportion of reliable variance associated with each factor The goodness of model-data fit for a three-factor solution was (subscale) after the variance associated with the single general better than the previous one-factor solution used as a reference factor has been partitioned out (68). The average variance (Table 4), explaining 55.8% of the total variance; moreover, extracted (AVE) was also estimated, defined as the amount loadings were congruent with the item theoretical provenance of variance captured by the construct vs. variance due to (Table 2). The uniqueness terms were lower than observed in the measurement error. Some authors suggest that the construct previous one-factor solution. The general simplicity values were has convergent validity if AVE ≥ 0.50, but values of ∼0.40 adequate (see Supplementary Material). The FDI (DE = 0.94; with a composite reliability > 0.60 are considered acceptable NJ = 0.97; NA = 0.94), construct replicability (DE: H = 0.89; (69). Discriminant validity between factors exists when the NJ: H = 0.93; NA: H = 0.88), and reliability (DE: ωS = 0.87; NJ: AVE values are greater than the squared correlation between ωS = 0.92; NA: ωS = 0.85) were appropriate. Inter-factor latent factors (69). The percentage of uncontaminated correlations correlations were “DE–NJ” ϕ = 0.56, “DE–NA” ϕ = 0.75, “NJ– (PUC) was estimated as the number of correlations between NA” ϕ = 0.58), and AVE was 0.50. Residual covariances were low, items from different factors divided by the total number of with average absolute values <0.001. correlations, which indicates the proportion of correlations The Schmid–Leiman approximation presented the same reflecting the possible general factor under study. When explained variance, uniqueness terms, and fit to the data as ECV and PUC values are >0.70, common variance can be the previous three-factor solution. Loadings in the general regarded as essentially unidimensional, but this is also true factor ranged from 0.45 (items NJ1 and NJ8) to 0.67 (item if the PUC is <0.80 but the ECV is >0.60 and the ωH is DE3), with positive weights in all of the corresponding >0.70 (61). theoretical provenance factors. The pure exploratory bifactor Frontiers in Psychiatry | www.frontiersin.org 5 November 2021 | Volume 12 | Article 659835 Frontiers in Psychiatry | www.frontiersin.org Soler et al. TABLE 2 | Descriptive statistics of the items and exploratory factor analyses (subsample 1). One factor Three factors Schmid–Leiman solution Exploratory bifactor Source Item Mean SD λ δ λ1 λ2 λ3 δ λ1 λ2 λ3 G δ λ1 λ2 λ3 G δ DE DE1 3.28 0.81 0.48 0.77 0.44 −0.04 0.16 0.70 0.23 −0.03 0.08 0.49 0.70 0.22 −0.09 0.01 0.52 0.68 DE2 3.44 0.92 0.57 0.68 0.44 0.10 0.12 0.64 0.23 0.08 0.06 0.54 0.64 0.24 0.06 0.01 0.53 0.64 DE3 3.23 0.93 0.68 0.54 0.63 0.06 0.12 0.43 0.33 0.04 0.06 0.67 0.43 0.40 0.05 −0.01 0.63 0.40 DE4 3.62 0.78 0.62 0.61 0.71 0.02 0.02 0.47 0.38 0.01 0.01 0.63 0.47 0.35 −0.06 −0.10 0.64 0.46 DE5 3.51 0.80 0.63 0.60 0.41 0.12 0.21 0.56 0.22 0.09 0.10 0.61 0.56 0.26 0.09 0.04 0.59 0.56 DE6 3.69 0.79 0.45 0.80 0.77 −0.11 −0.12 0.60 0.41 −0.08 −0.06 0.48 0.60 0.48 −0.08 −0.14 0.43 0.58 DE7 3.43 1.06 0.51 0.74 0.45 −0.01 0.15 0.67 0.24 −0.01 0.07 0.51 0.67 0.32 0.01 0.04 0.47 0.67 DE8 3.62 1.01 0.49 0.76 0.91 −0.07 −0.25 0.49 0.49 −0.05 −0.12 0.51 0.49 0.53 −0.07 −0.23 0.46 0.47 DE9 3.84 0.86 0.65 0.58 0.66 0.04 0.07 0.46 0.35 0.03 0.03 0.65 0.46 0.39 0.01 −0.05 0.62 0.46 NJ NJ1 3.22 1.10 0.61 0.63 0.16 0.62 −0.10 0.57 0.08 0.47 −0.05 0.45 0.57 −0.49 0.13 −0.35 0.80 0.02 NJ2 3.68 0.96 0.74 0.45 0.12 0.69 0.03 0.39 0.07 0.52 0.01 0.58 0.39 −0.15 0.44 −0.10 0.66 0.37 NJ3 4.05 0.97 0.70 0.51 −0.10 0.80 0.10 0.35 −0.05 0.61 0.05 0.53 0.35 −0.15 0.65 0.01 0.52 0.34 NJ4 3.43 1.06 0.65 0.57 0.01 0.76 −0.02 0.45 0.01 0.57 −0.01 0.48 0.45 −0.18 0.54 −0.09 0.52 0.45 NJ5 3.88 0.97 0.74 0.45 0.04 0.83 −0.02 0.29 0.02 0.62 −0.01 0.56 0.29 −0.09 0.67 −0.07 0.55 0.28 NJ6 4.05 0.91 0.71 0.50 −0.09 0.91 −0.01 0.27 −0.05 0.68 −0.01 0.51 0.27 −0.20 0.71 −0.07 0.52 0.25 NJ7 3.78 1.00 0.65 0.58 0.07 0.65 0.02 0.50 0.04 0.49 0.01 0.50 0.50 0.01 0.56 −0.03 0.46 0.47 NJ8 3.56 1.15 0.63 0.60 0.01 0.80 −0.10 0.43 0.01 0.61 −0.04 0.45 0.43 −0.25 0.53 −0.15 0.53 0.43 6 NA NA1 4.52 1.42 0.65 0.58 0.02 0.14 0.60 0.51 0.01 0.11 0.28 0.64 0.51 0.11 0.14 0.28 0.62 0.51 NA2 5.22 1.04 0.54 0.71 −0.01 −0.10 0.75 0.52 −0.01 −0.08 0.35 0.60 0.52 0.21 −0.01 0.39 0.54 0.52 NA3 4.93 1.29 0.62 0.62 0.14 −0.02 0.61 0.49 0.07 −0.02 0.29 0.65 0.49 0.23 0.03 0.30 0.60 0.49 NA4 4.94 1.26 0.54 0.71 −0.08 −0.03 0.75 0.54 −0.04 −0.02 0.35 0.58 0.54 0.15 0.06 0.39 0.53 0.54 NA5 5.12 1.07 0.41 0.83 0.14 −0.18 0.55 0.68 0.07 −0.14 0.26 0.48 0.68 0.23 −0.10 0.27 0.44 0.68 NA6 4.91 1.29 0.57 0.67 −0.12 0.01 0.80 0.49 −0.06 0.01 0.37 0.61 0.49 0.09 0.05 0.40 0.58 0.49 NA7 4.96 1.10 0.57 0.67 −0.01 0.07 0.62 0.58 −0.01 0.05 0.29 0.58 0.58 0.16 0.12 0.31 0.52 0.58 SD, standard deviation; Characteristics of the matrix: determinant <0.001; KMO = 0.94; Bartlett’s statistic = 3,702.40 (df = 276; p < 0.001): Mardia’s statistic = 37.71 (p < 0.001). λ, factorial weight; δ, uniqueness; G, general factor; DE, Decentering; NJ, Non-Judging from the Five Facets of Mindfulness Questionnaire. NA, Non-Attachment. Decentering, Acceptance, and Non-Attachment November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment model explained 59.6% of the total variance and improved the solution had a poor fit to the data (Table 4), as in the previous fit to the data (Table 4). The uniqueness terms of the DE and NJ EFA, and thus, it was discarded. items were decreased, while the NA terms were exactly the same The three-correlated factor solution explained 58.4% of total (Table 2). The general factor (G) presented loadings between 0.43 variance. It showed adequate loadings (Table 5; all significant: (item DE6) and 0.80 (NJ1); the AVE was 0.53. The FDI values p < 0.001), ranging from 0.51 (item NA5) to 0.87 (item NJ5). were appropriate (DE = 0.92; NJ = 0.92; NA = 0.84; G = 0.98). The fit to the data was acceptable (Table 4). The FDI values The construct replicability of DE (H = 0.85), NJ (H = 0.84) were good (DE = 0.94; NJ = 0.97; NA = 0.93). Construct and the general factor (H = 0.96) were good, but the value of replicability and reliability estimates were appropriate (DE: H the NA subscale was fair (H = 0.71). The omega hierarchical of = 0.88, ωS = 0.87; NJ: H = 0.94, ωS = 0.94; NA: H = 0.96, the general factor was good (ωH = 0.80), but the values of the ωS = 0.85). Inter-factor latent correlations were as follows: subscales were low (DE: ωHS = 0.26; NJ: ωHS = 0.43; NA: ωHS = “DE–NJ” ϕ = 0.58, “DE–NA” ϕ = 0.80, “NJ–NA” ϕ = 0.57; 0.23). Residual covariances were very low, with average absolute the AVE was 0.52. Residual co-variances were low and equally values <0.001. distributed among all the items, with an average absolute value of 0.04. Confirmatory Factor Analyses The single second-order factor solution presented the same Descriptive statistics for all items (subsample 2) are shown in fit, factor loadings, and uniqueness terms as that obtained in Table 5. The BIC dimensionality test in subsample 2 also resulted the three-correlated factor solution, with high and significant in a one-factor solution (Table 3) that explained 41.1% of the total second-order loadings [DE (γ = 0.90, R2 = 0.81, p < 0.001), variance, with significant loadings in all cases (Table 5), ranging NJ (γ = 0.64, R2 = 0.41, p < 0.001), NA (γ = 0.89, R2 = 0.79, from 0.37 (item NA5) to 0.80 (item NJ3). However, the one-factor p < 0.001)], and a total omega value of ω = 0.95. The bifactor model improved the fit to the data (Table 4), and the general factor (G) showed positive and significant loadings in all the items (Table 5), ranging from 0.35 (item NA5) to 0.77 (DE5). The TABLE 3 | BIC dimensionality tests. averaged variance extracted (AVE = 0.54), construct replicability (H = 0.93), factor determinacy (FDI = 0.89), explained common Subsample 1 Subsample 2 variance (ECV = 0.61), and omega hierarchical (ωH = 0.79) Factors BIC Factors BIC of the general factor were appropriate, with a PUC of 0.69. By contrast, the partialized subscales were weak in terms of construct 0 13,804.41 0 15,113.58 replicability and factor determinacy, except for NJ (DE: H = 0.48, 1 8,459.02* 1 8,368.37* FDI = 0.32; NJ: H = 0.86, FDI = 0.77; NA: H = 0.52, FDI = 2 9,053.39 2 9,196.81 0.49). The omega hierarchical values of the subscales were also 3 10,303.34 3 10,778.03 insufficient (DE: ωHS = 0.11; NJ: ωHS = 0.59; NA: ωHS = 0.24). *Recommended number of common factors, 1; BIC, Schwartz’s Bayesian Residual covariances were low and equally distributed among all information criterion. the items, with an average absolute value of 0.02. TABLE 4 | Fit indexes for the EFA, CFA, and SEM analyses. Group/Model χ2 df χ2 /df CFI TLI RMSEA (90% CI) SRMR EFA: subsample 1 One-factor 1,233.71* 252 4.90 0.946 0.941 0.092 (0.086–0.098) 0.099 Three correlated factors 377.56* 207 1.82 0.985 0.980 0.053 (0.046–0.060) 0.033 Schmid-Leiman 377.56* 207 1.82 0.985 0.980 0.053 (0.046–0.060) 0.033 Exploratory bifactor 302.45* 186 1.63 0.986 0.980 0.053 (0.046–0.060) 0.030 CFA: subsample 2 One-factor 1,386.54* 252 5.50 0.713 0.686 0.122 (0.116–0.128) 0.110 Three correlated factors 432.29* 249 1.74 0.954 0.949 0.049 (0.041–0.057) 0.045 One second-order factor 432.29* 249 1.74 0.954 0.949 0.049 (0.041–0.057) 0.045 Bifactor model 362.16* 228 1.59 0.966 0.959 0.044 (0.035–0.052) 0.041 Total sample Bifactor model 439.25* 228 1.93 0.971 0.965 0.039 (0.034–0.045) 0.029 Structural model (SEM) 1,509.97* 752 2.01 0.944 0.938 0.041 (0.038–0.044) 0.046 χ 2 , minimum value of the discrepancy; df, degrees of freedom; CFI, comparative fit index; TLI, Tucker-Lewis index; RMSEA (90% CI), root mean square error of approximation (90% confidence interval); SRMR, standardized root mean square residual. *p < 0.001; EFA, exploratory factor analyses; CFA, confirmatory factor analyses; SEM, structural equation modeling (The graphical representation of SEM is in Figure 1). Frontiers in Psychiatry | www.frontiersin.org 7 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment TABLE 5 | Descriptive statistics of the items and confirmatory factor analyses (subsample 2). One factor Three factors† Bifactor model Source Item Mean SD λ δ λ specific δ λ specific G δ EQ DE1 3.30 0.83 0.50*** 0.75 0.62*** 0.62 0.18* 0.58*** 0.63 DE2 3.42 0.90 0.57*** 0.68 0.66*** 0.56 −0.01 0.69*** 0.52 DE3 3.18 0.88 0.50*** 0.75 0.69*** 0.52 0.19** 0.65*** 0.55 DE4 3.53 0.86 0.49*** 0.76 0.64*** 0.59 0.26*** 0.58*** 0.60 DE5 3.47 0.83 0.61*** 0.63 0.71*** 0.50 −0.06 0.77*** 0.40 DE6 3.65 0.74 0.46*** 0.79 0.54*** 0.71 0.32*** 0.47*** 0.68 DE7 3.37 1.05 0.39*** 0.85 0.56*** 0.69 0.20** 0.52*** 0.69 DE8 3.51 0.96 0.54*** 0.71 0.68*** 0.54 0.51*** 0.57*** 0.42 DE9 3.78 0.77 0.60*** 0.64 0.79*** 0.38 0.45*** 0.70*** 0.31 NJ NJ1 3.21 1.07 0.67*** 0.55 0.73*** 0.47 0.62*** 0.40*** 0.37 NJ2 3.68 1.01 0.75*** 0.44 0.80*** 0.36 0.63*** 0.48*** 0.31 NJ3 3.97 1.03 0.80*** 0.36 0.83*** 0.31 0.62*** 0.55*** 0.37 NJ4 3.39 1.08 0.76*** 0.42 0.80*** 0.36 0.62*** 0.50*** 0.23 NJ5 3.80 0.99 0.79*** 0.38 0.87*** 0.24 0.74*** 0.47*** 0.30 NJ6 3.87 0.98 0.79*** 0.38 0.84*** 0.29 0.66*** 0.52*** 0.34 NJ7 3.77 1.01 0.78*** 0.39 0.81*** 0.34 0.62*** 0.52*** 0.29 NJ8 3.54 1.11 0.79*** 0.38 0.84*** 0.29 0.66*** 0.53*** 0.46 NAS NA1 4.58 1.35 0.56*** 0.69 0.75*** 0.44 0.29*** 0.68*** 0.45 NA2 5.13 1.12 0.52*** 0.73 0.65*** 0.58 0.30*** 0.57*** 0.59 NA3 4.94 1.25 0.53*** 0.72 0.74*** 0.45 0.32*** 0.66*** 0.46 NA4 4.80 1.25 0.61*** 0.63 0.73*** 0.47 0.37*** 0.62*** 0.48 NA5 4.96 1.10 0.37*** 0.86 0.51*** 0.74 0.47*** 0.35*** 0.65 NA6 4.81 1.17 0.52*** 0.73 0.66*** 0.56 0.43*** 0.53*** 0.53 NA7 4.82 1.08 0.54*** 0.71 0.67*** 0.55 0.31*** 0.59*** 0.56 SD, standard deviation; Characteristics of the matrix, determinant <0.001; KMO = 0.95; Bartlett’s statistic = 4,108.80 (df = 276; p < 0.001), Mardia’s statistic = 31.98 (p < 0.001); λ, loading; δ, uniqueness; G, general factor; EQ, experiences questionnaire; NJ, non-judging (from the FFMQ); NAS, non-attachment scale. † The one second-order model showed the same specific loadings and uniqueness terms as the three-factor model, *p < 0.001, ** p < 0.005, *** p < 0.05. Nomological Validity were low and equally distributed among all the items, with a Given the appropriateness of the general factor (G) resulting mean absolute value of 0.03. from the bifactor model, we tested the nomological validity of this factor for resilience and depression in the full sample (n = 608). As shown in Table 4, the measurement bifactor model of the new DISCUSSION construct fit the data well in the total sample. In addition, the fit indexes of the structural nomological model were acceptable, The main objective of this study was to test the hypothesis with the following values: χ 2 = 1,509.97; df = 752 (p < 0.001); that there is an underlying general factor beyond the constructs χ 2 /df = 2.01; CFI = 0.944; TLI = 0.938; RMSEA (90% CI) = of acceptance, decentering, and non-attachment. Despite the 0.041 (0.038–0.044); and SRMR = 0.046. The raw inter-factor apparent semantic differences among them, they have shown latent correlations between the constructs showed the following to be psychometrically closely related (31, 70). Our findings values: “G–resilience” ϕ = 0.81, p < 0.001; “G–depression” ϕ = suggest that a more general factor that subsumes all three of these −0.66, p < 0.001; “resilience–depression” ϕ = −0.58, p < 0.001). components appears to exist. Considering the content of the The results of the structural model are represented in Figure 1. As referred domains and the theoretical connection among them, that figure shows, all the factorial loadings of the general factor suggested by contemplative authors (71–73), we decided to label (G) resulting from the bifactor model, as well as the factorial this general factor “Delusion of Me” (DoM) (Figure 1). loadings of the other constructs (e.g., resilience and depression), This idea of a general factor was explored and confirmed were significant. The explanatory power of the general factor through various analytical models applied to two distinct, (G) on resilience and depression were, respectively, R2 = 0.65 randomly selected subsamples. Considering the three correlated, (β = 0.81, p < 0.001) and R2 = 0.44 (β = −0.55, p < 0.001). first-order factor model and following the criteria proposed by The adjusted link between resilience–depression showed a non- Fornell and Larcker (69), we established convergent validity significant value (β = −0.13, p = 0.070). Residual co-variances among the original components (non-judgment, decentering, Frontiers in Psychiatry | www.frontiersin.org 8 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment FIGURE 1 | Nomological validity of the “Delusion of Me” (DoM) general factor regarding resilience and depression. Structural equations model of the nomological validity of the “Delusion of Me” (DoM) general factor regarding the constructs of “resilience” and “depression.” The circles represent latent components, and the rectangles are observable variables (items). One-way arrows represent factor loadings (in the measurement model) or standardized slopes (in the structural model). Dashed lines represent non-significant values in the corresponding factor loadings or standardized slopes (i.e., p > 0.05). *p < 0.001. and non-attachment). However, discriminant validity was only of non-judging, decentering, and non-attachment) suggests observed between decentering and non-judgment, and between plausible alternative hypotheses. For instance, the best fit of the non-attachment and non-judgment; in other words, decentering bifactor model does not deny the observation of an excellent and non-attachment appear to be very closely related facets. The fit for the three-factor structure with a second-order factor of bifactor approximation was the model that best fit the data. Based DoM (that can be obtained as an aggregate of the scores of on this model, the ECV—given the context of the PUC and ωH the three factors). Furthermore, in the bifactor model, items values obtained—suggested the one-dimensional nature of the of all the three factors loaded significantly on the respective pool of items, with adequate loadings on the general factor and latent factors (except two items of decentering) even after appropriate fit indexes (61). Simplicity and factor determinacy fragmenting the general factor of the DoM. Therefore, this also were good, and the factor score estimates unambiguously may be considered to support the notion that the constructs reflected the latent levels of the general factor they attempted of non-judging, decentering, and non-attachment, even though to estimate (62). The AVE, ωH , and the H index were also correlating with each other and jointly representing the higher appropriate, indicating that the general factor was well-defined order constructs of the DoM, also represent constructs closely (61). However, the original theoretical subscales (e.g., non- related to mindfulness itself. judgment, decentering, and non-attachment) remained marginal The term “delusion” is related to the lack of decentering. in practically all psychometric indexes after adjusting for the Cognitive fusion refers to a mental process in which verbal general factor. This suggests that the unique characteristics of events (i.e., thoughts) prevail over direct experience in terms of each may be minimal when compared with the shared common influence on behavior (46). As a result, the individual begins aspects present in the general factor that seems to group them. to respond to mental constructs as if they were physical facts We have seen that several statistical indices support the (25, 74). On the contrary, decentering enables the individual to let unidimensional construct represented by the item pools of the go of needless entanglement with private experiences (46) such self-report measures of non-judging, decentering, and non- are thoughts and emotions. Decentering includes three separate attachment. However, failure to obtain an excellent fit for a single components: meta-awareness of subjective experience, reduced factor model and equally excellent fit for three correlated factors reactivity to thought content, and dis-identification from internal and a second-order factor (linked with the first-order factors experience (75). Decentering allows us to step back from an Frontiers in Psychiatry | www.frontiersin.org 9 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment experience to examine it as separate from the self and to facilitate the self-narrative is positive, attachment arises, and people dis-identification from internal experience. In Buddhism and may overestimate themselves and inadvertently cause harm to contextual sciences this conceptual self is an illusion (46, 73, 76, themselves and others. 77). People describe themselves in terms of their roles, history, The DoM index enables us to measure the degree to which attributes, and dispositions that blended all together constitute expectancy of a certain outcome is confounded with actual facts. our narrative self that, through fusion, became “facts” (46). More precisely, the degree to which preferences are converted Although illusory, this entity has intentionality, an interest in into desires of influence and become strong demands to “the obtaining certain results and not others (78). This fundamental reality,” in other words, the overestimation of one’s own capacity mismatch between this simplified way of perceiving the world, to alter the events. based on personal preferences, and the true nature of reality is In this sense, DoM index is also inversely related to the what Buddhism calls “ignorance” (72). Ignorance refers to the concept of wei-wu-wei, the Taoist notion whereby an action illusion that—beyond our preferences, impressions, or mental is performed without an intention, and the mental process events—there is something that constitutes a separate entity, is not a means to achieve any end (53), or the concept of which could be called a self (78). This “mistakenly reified self- “reflexive awareness” from Buddhism where the phenomenal image” (44) emerges out of the “mental simulation” conforming form of subjectivity is presented in an intransitive way (85). The the self. DoM index is also coherent with the idea of a “false sense of The closest related psychological term to DoM is the self-as-a- authorship” in the Vedanta tradition in which the belief that we content, a construct pertaining to Acceptance and Commitment are in control of our will is a delusion (85–87). Finally, from Therapy (46), in which any private experience or content the contextual behavioral science perspective also, there is no becomes the self by fusion or identification. In other words, our agency, and every thought, as any other behavior, is defined self becomes our current particular content of mind. When I say by a limited number of variables (88). For instance, the self in “my body,” the self becomes the owner of the body. The self is our Skinner’s approach has to do with responding under the control vision when “I see. . . ” and my thought when I say, “I think. . . ” of the environment (77).The data obtained in the present study (79). When an individual identifies with a particular self-concept indicate that DoM might have a powerful influence on suffering (i.e., I’m the kind of person who . . . ) alternative views are less (considering depression as an index), even when compared with likely to be seen and inconsistencies are threatening (39, 46). resilience. Nevertheless, additional research is needed to further Interestingly, narrative self that could be considered a by- explore the DoM role on psychopathology. product of self-as-content, has been linked to medial prefrontal The DoM approach offers two significant advantages over cortex (MPFC) activity while the experiential self, momentary other approaches: first, as its referents, its potential as a self-reference centered on the present without selecting any transdiagnostic construct. We performed a preliminary test to sensory object, with pervasive reductions in the same area determine the association between the DoM index and a common and increased engagement of a right lateralized network (44). clinical index (e.g., depressive symptoms). As expected, self- MPFC, among other areas, and the left hemisphere dominance reported symptoms of depression were inversely and significantly on “autobiographical-self ” vs. non-verbal bodily processes or related to DoM. This finding is congruent with previous studies “core self ” was also pointed out in later research, also targeting that have consistently reported inverse correlations between sources of self-reference (80). On the other hand, decentering acceptance, decentering, and non-attachment with depression from default mode of egocentricity was associated with greater indexes in both clinical and non-clinical populations (1, 2, 28, response within the posterior insula, supramarginal gyrus, and 29, 89). Nevertheless, future research testing the relation between ventrolateral prefrontal cortex, with a tendency toward greater the DoM and the existent psychopathology related to depression right hemisphere involvement (81). is needed to draw the potential transdiagnostic nature and utility The essence of attachment relies on grasping or clinging to of the construct. things, others, or even our own self (73). Sahdra et al. (29, 31) As we hypothesized, there was also a strong positive and stated that attachment requires being stuck or fixated on ideas, significant association between the DoM index and resilience. images, or sensory objects, and feeling an internal pressure to Resilience is a broad construct (with no single definition) acquire, hold, avoid, or change them. By contrast, as previously that includes various aspects of psychological resistance (90). stated, acceptance could be defined as “experiencing events fully Resilience could be described as a multidimensional factor that and without defense, as they are” (82). This also implies that moderates negative emotions and distress, thereby facilitating the individual does not try to change, avoid, or escape from a a flexible adaptation to suboptimal conditions (91). There is given experience (74, 83). Along these lines, the idea of grasping, substantial empirical evidence that psychological resilience can clinging, or “feeling an internal pressure to” is similar to the help an individual to regain or maintain physical and mental absence of acceptance and similar to the notion of working either health (92–94). Interestingly, when resilience, depression, and against or in favor of a particular outcome. Both non-acceptance DoM were considered together (Figure 1), resilience lost its and attachment have in common the transition from preference explanatory power over depression symptom scores and was to need through fusion. submerged by the power of DoM to explain depressive mood. Most people are attached to their constructs of self. This is It should be noted that all three DoM constituents have been the origin of a myriad of problems (84). If their self-narrative considered relevant in “third wave” clinical approaches. For happens to be negative, it feeds aversion and avoidance. If instance, in Acceptance and Commitment Therapy, flexible and Frontiers in Psychiatry | www.frontiersin.org 10 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment healthy psychological functioning relies on the six corners of the Tolle (104) pointed out, when “my toy” is broken or taken away, hexaflex model (i.e., flexible attention to the present moment, intense suffering arises, not because of the intrinsic value of the acceptance, defusion, self as a context, values, and committed toy, but because of the thought of “mine.” The toy becomes part action) two of which are contained in DoM (46). Along of the child’s developing sense of self, of “I,” and, as we see, the with ACT, therapies such MBCT, mindfulness-based relapse pronoun “mine” is inseparable of the attachment quality. Finally, prevention (MBRP) and DBT exactly or similar terms such as attachment and lack of acceptance seem to be the heads and tails non-reactivity, non-judgmental, aversion, craving, acceptance, of the same phenomenon, and have the outcomes that “I” want. willingness, “letting go,” decentering, or defusion are defined and Recently, it has been proposed the term “hypo-egoic” to specifically targeted within the therapy process. Congruently, the describe states where (1) conceptual self-awareness is low; (2) three DoM constituents have been studied as mechanisms of the phenomenal self is not highly individuated from its context; change for “third wave” clinical approaches when targeting a and (3) the person is not selfishly invested in the outcome wide diversity of mental disorders such as depression, anxiety, of a particular situation (105). The usual egoic form of self substance use disorders, and personality disorders (95–98). contributes to depression, anxiety, anger, jealousy, and other In this regard, our findings suggest that, by training one negative emotions that we experience. It enables us to ruminate of these facets, the individual would actually train all three. If and, interpersonally, undermine our relationships, and it can be this is true, it could help to tailor the individual’s treatment a source of social conflict by promoting dislike for those who are according to their specific needs and preferences. For example, different from us (106). a given individual may be reluctant to practice mindfulness In summary, DoM index is psychometrically robust and of thoughts to promote decentering, but could be more open could be a reliable measure in mindfulness field. The construct to psycho-education to foster acceptance, or to self-distancing may function as a transdiagnostic index to measure the training (89) and track the impermanence of experiences (90) overestimation of one’s own capacity to alter the events. DoM to increase defusion or to random acts of kindness (99) to appears to be powerful connected to both negative affect encourage non-attachment. In this regard, future studies should and resilience. Additionally, and beyond the strictly individual analyze the utility and the potential therapeutic value of this assessment, the construct has the potential to be useful in new construct. As Fahlberg and Fahlberg (100) suggest, if relation to prosocial and lifestyle variables and in the context of we could only weaken the hold of our conditioned thoughts, interventions designed to induce a “hypo-egoic” effect. emotions, and behaviors this could lead to a more pro-social and rational life and therefore mitigate the negative social and environmental consequences of egoic behavior and lifestyles LIMITATIONS AND FUTURE DIRECTIONS (101). The second strength of the DoM index is that it appears to be psychometrically tenable. As we stated in the The main limitation of the present study is that the results are introduction, the currently available measures of mindfulness based only on a limited number of instruments, despite the wide are inconsistent and poorly correlated. In a recent review, variety of scales that could have been used to assess these three Park et al. (14) concluded that “. . . the current mindfulness constructs. Moreover, other constructs such as equanimity or scales have important conceptual differences and none can be compassion could have been included in the model, as these have clearly recommended based on their psychometric properties. indirect implications for self-concept and self-reference (79). In Researchers should proceed with caution before optimizing addition, all of the measures used in this study were self-reported, any mindfulness intervention based on existing scales.” In this with the inherent potential for bias such as social desirability or regard, approximately half of the studies involving mindfulness- acquiescence. In particular, the possibility of “common method based interventions either fail to show any significant pre-post bias” should not be discarded, as using the same methodology treatment effect in self-reported measures of mindfulness, or the to assess different constructs often loads on a single factor due outcomes are no better than those obtained with other active to the similarity of the assessment, and not because of content interventions (102), pointing to weaknesses in the tools used to overlap between the constructs. The participant recruitment and measure the effects of mindfulness practice. We consider DoM as methodology of survey over regional online websites and social a trait measure, since EQ, NAS, and FFMQ fail on this category. media could have also hamper the generalizability of our results. Similarly, we also think on DoM, mainly, as a by-product of Another limitation is that, although we assessed the association mindfulness practice since Decentering and Non-attachment can between DoM and depression, our sample did not include clinical be considered as consequences of mindfulness (27, 70). On diagnosis, and therefore, we do not know the clinical applicability the other hand, from a Western point of view of mindfulness, of our findings. In addition, the study design does not allow acceptance is part of it, rather than an aftermath (7). Although us to make causal inferences between variables, and also the our study does not include any pre-post assessment, we think that role of possible confounders was not evaluated. Nevertheless, it is highly likely that the DoM index will be sensitive to change it is worth highlighting that one important strength was the given the strong sensitivity of its components (i.e., EQ, NAS, and use of two randomly selected sub-samples to cross-validate Non-Judgment from FFMQ) (26, 27, 103). results obtained. DoM intends to be useful in assessing interventions targeting Future research must consider other outcomes related to “ego.” The identification with thoughts that represent things and this construct, ideally including ecological measurement models ideas transfer the sense of identity from them to our self. As of mindfulness in daily life. Additionally, the utility of the Frontiers in Psychiatry | www.frontiersin.org 11 November 2021 | Volume 12 | Article 659835 Soler et al. Decentering, Acceptance, and Non-Attachment DoM concept in clinical samples should be tested in future AUTHOR CONTRIBUTIONS research, especially through longitudinal studies to test causal inferences. In summary, DoM emerges as a robust, potentially JS, AC, and JG-C designed the study. ED-C and SG executed transdiagnostic concept, with possible influences on well- the fieldwork. JM-M performed the data analyses. JS and JM-M being, but further studies should analyze its utility at the were in charge of writing the manuscript and responding to therapeutic level. reviewers comments. JG-C, AC, and BA assisted in the editing of the manuscript. All authors read and approved the final version CONCLUSIONS of the manuscript. The study shows that there is a general factor that subsumes the FUNDING constructs of acceptance, decentering, and non-attachment. We used analytical models applied to two distinct, randomly selected JM-M was supported by the Wellcome Trust Grant subsamples, but despite the apparent semantic differences among (104908/Z/14/Z). Additionally, Ciberobn is an initiate of them, they have shown to be psychometrically closely related. the ISCIII. Thus, we have labeled it as “Delusion of Me” (DoM) to refer to the delusion of control of our will, of our “self.” This construct ACKNOWLEDGMENTS appears to correlate negatively to psychopathology and positively to resilience. This research was funded in whole, or in part, by the Wellcome Trust [104908/Z/14/Z], by the Master in Mindfulness DATA AVAILABILITY STATEMENT Programme and the Chair of Contemplative Sciences, University of Zaragoza (Zaragoza, Spain), and the Instituto de Investigación The original contributions presented in the study are included Sanitaria Aragón (IIS Aragon; Zaragoza, Spain). The funding in the article/Supplementary Material, further inquiries can be source had no influence on the design of the study, data collection directed to the corresponding authors. and analysis, or the writing of the manuscript. For the purpose of open access, the authors have applied a CC BY public copyright ETHICS STATEMENT license to any Author Accepted Manuscript version arising from this submission. The study protocol was approved by the corresponding regional health authority, the Aragon Ethics Committee (CEICA), Spain SUPPLEMENTARY MATERIAL (registry: PI12/00083) and all participants signed a consent form indicating their willingness to participate. Participants were The Supplementary Material for this article can be found informed about the purpose of the study and informed that all online at: https://www.frontiersin.org/articles/10.3389/fpsyt. response would be treated confidentially. 2021.659835/full#supplementary-material REFERENCES 7. 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