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Factor Structure of the Five Facets Mindfulness Questionnaire (FFMQ) (15 Items) in a Collectivist Society--Pakistan

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Title: Factor Structure of the Five Facets Mindfulness Questionnaire (FFMQ) (15 Items) in a Collectivist Society--Pakistan
Language: English
Authors: Iqbal, Farhan (ORCID 0000-0002-6227-1979), Iqbal, Farah, Humayun, Ghazal Khawaja
Source: Psychology in the Schools. Jul 2023 60(7):2502-2519.
Availability: Wiley. Available from: John Wiley & Sons, Inc. 111 River Street, Hoboken, NJ 07030. Tel: 800-835-6770; e-mail: cs-journals@wiley.com; Web site: https://www.wiley.com/en-us
Peer Reviewed: Y
Page Count: 18
Publication Date: 2023
Document Type: Journal Articles
Reports - Research
Education Level: Higher Education
Postsecondary Education
Descriptors: Questionnaires, Metacognition, Personality Measures, Foreign Countries, Factor Structure, Factor Analysis, College Students, Collectivism, Correlation, Multivariate Analysis, Scores, Computation
Geographic Terms: Pakistan
Assessment and Survey Identifiers: NEO Five Factor Inventory
DOI: 10.1002/pits.22875
ISSN: 0033-3085
1520-6807
Abstract: Despite several theoretical, structural, and statistical issues reported against Five Facets Mindfulness Questionnaire (FFMQ), most studies in the educational sector of Pakistan use it without analyzing its factor structure. Since culture might change the structure, this first systematic study filled the gap and explored the factor structure of the least researched FFMQ (15 items) among nonmeditator university students in collectivist society Pakistan while employing optimal statistical techniques. We interpreted our results culturally and also conducted a cross-cultural examination with the Spanish sample. We employed polychoric data to conduct exploratory factor analysis (EFA) and confirmatory factor analysis (CFA). The EFA extracted a four-factor (10 items) solution without the Describe facet and showed good inter-item correlation. The CFA exhibited a theory-based two-factor hierarchical model (self-regulation, orient to experience) with good fit indices, acceptable factor loadings, correct directions, and strong reliabilities (coefficient H > 0.8). The multivariate analysis showed no difference in mindfulness between males and females, except for Nonreact. The first multigroup CFA (MGCFA) showed metric invariance while examining the polarity of items across gender. In cross-cultural multivariate analysis, we found no difference in mindfulness between Pakistan and Spanish samples, except for Nonjudge, but the second MGCFA showed metric invariance in the correlated model only. Our data suggest that the two-factor hierarchical structure, without Describe facet, provides a more parsimonious and culturally accepted structure among nonmeditators in Pakistan. However, due to the independent nature of the two factors, the total score cannot be computed. The FFMQ (15 items) appears individualistic, and including culturally acceptable items will improve its universality.
Abstractor: As Provided
Entry Date: 2023
Accession Number: EJ1379649
Database: ERIC
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  Value: <anid>AN0164066401;pis01jul.23;2023Jun05.08:09;v2.2.500</anid> <title id="AN0164066401-1">Factor structure of the Five Facets Mindfulness Questionnaire (FFMQ) (15 items) in a collectivist society—Pakistan </title> <p>Despite several theoretical, structural, and statistical issues reported against Five Facets Mindfulness Questionnaire (FFMQ), most studies in the educational sector of Pakistan use it without analyzing its factor structure. Since culture might change the structure, this first systematic study filled the gap and explored the factor structure of the least researched FFMQ (15 items) among nonmeditator university students in collectivist society Pakistan while employing optimal statistical techniques. We interpreted our results culturally and also conducted a cross‐cultural examination with the Spanish sample. We employed polychoric data to conduct exploratory factor analysis (EFA) and confirmatory factor analysis (CFA). The EFA extracted a four‐factor (10 items) solution without the Describe facet and showed good inter‐item correlation. The CFA exhibited a theory‐based two‐factor hierarchical model (self‐regulation, orient to experience) with good fit indices, acceptable factor loadings, correct directions, and strong reliabilities (coefficient H > 0.8). The multivariate analysis showed no difference in mindfulness between males and females, except for Nonreact. The first multigroup CFA (MGCFA) showed metric invariance while examining the polarity of items across gender. In cross‐cultural multivariate analysis, we found no difference in mindfulness between Pakistan and Spanish samples, except for Nonjudge, but the second MGCFA showed metric invariance in the correlated model only. Our data suggest that the two‐factor hierarchical structure, without Describe facet, provides a more parsimonious and culturally accepted structure among nonmeditators in Pakistan. However, due to the independent nature of the two factors, the total score cannot be computed. The FFMQ (15 items) appears individualistic, and including culturally acceptable items will improve its universality.</p> <p>Practitioner points: The Describe facet of FFMQ (15 items) does not appear valid in a collectivist society. Caution may be exercised in any evaluation.The two‐factor hierarchical model is better suited than a single mindfulness factor in a collectivist society.The computation of separate scores for the two higher‐order factors is preferred over the total score.</p> <p>Keywords: cross‐cultural; factor structure; FFMQ (15 items); mindfulness; Pakistan</p> <hd id="AN0164066401-2">INTRODUCTION</hd> <p>Mindfulness, most popularly defined as present‐moment awareness, is widespread. Scholars are analyzing it in as diverse fields as criminal justice (Auty, [<reflink idref="bib2" id="ref1">2</reflink>]), clinical practices (Gupta, [<reflink idref="bib44" id="ref2">44</reflink>]), and the corporate sector (Bayle‐Cordier et al., [<reflink idref="bib10" id="ref3">10</reflink>]). Fortunately, the educational sector is not far behind. Powietrzyńska and Noble ([<reflink idref="bib77" id="ref4">77</reflink>]) consider mindfulness an antidote to current education's inhuman and individualistic nature. The extant literature highlights various benefits of mindfulness for teachers and students alike, including the management of stress, mental health, academic achievement, and working memory (Caballero et al., [<reflink idref="bib21" id="ref5">21</reflink>]; Mrazek et al., [<reflink idref="bib73" id="ref6">73</reflink>]; Von der Embse et al., [<reflink idref="bib95" id="ref7">95</reflink>]; Zarate et al., [<reflink idref="bib102" id="ref8">102</reflink>]). In line with the western world, there is a burgeoning interest among scholars in Pakistan to examine the relationship of mindfulness with other constructs in the educational sector (Iqbal et al., [<reflink idref="bib54" id="ref9">54</reflink>]; Malik & Perveen, [<reflink idref="bib68" id="ref10">68</reflink>]; Sibghat‐ullah & Batool, [<reflink idref="bib83" id="ref11">83</reflink>]; Zahra & Riaz, [<reflink idref="bib100" id="ref12">100</reflink>]). Though various scales exist to measure mindfulness, researchers primarily used a widely reported Five Facets Mindfulness Questionnaire (FFMQ; Baer et al., [<reflink idref="bib7" id="ref13">7</reflink>]) because it captured multiple dimensions of mindfulness. However, to the authors' knowledge, no systematic study has yet probed how FFMQ, especially the least investigated FFMQ (15 items), might function in Pakistan. Since culture might change the structure of any scale (Henrich et al., [<reflink idref="bib48" id="ref14">48</reflink>]), it is essential to investigate the factor structure of all scales before they are used in different cultures to prevent inadvertent interpretation of results in applied settings, including the educational sector. In this first investigative study, we intended to examine the factor structure of the least investigated FFMQ (15 items) among university students in Pakistan using optimal techniques.</p> <p>Baer et al. ([<reflink idref="bib7" id="ref15">7</reflink>]) conducted exploratory factor analysis (EFA) while combining statements from the existing five different scales, namely, the Mindfulness Attention and Awareness Scale (Brown & Ryan, [<reflink idref="bib17" id="ref16">17</reflink>]), the Freiburg Mindfulness Inventory (Buchheld et al., [<reflink idref="bib18" id="ref17">18</reflink>]), the Kentucky Inventory of Mindfulness Skills (KIMS; Baer et al., [<reflink idref="bib6" id="ref18">6</reflink>]), the Cognitive and Affective Mindfulness Scale (Feldman et al., [<reflink idref="bib30" id="ref19">30</reflink>]), and the Mindfulness Questionnaire (Chadwick et al., [<reflink idref="bib23" id="ref20">23</reflink>]) to extract 39 items for FFMQ. The confirmatory factor analysis (CFA) revealed mindfulness as a hierarchical concept underpinned by five first‐order latent variables: Observe (noticing or attending to the internal experiences), Describe (labeling internal experiences externally with words), Actaware (attending to one's activities in the present moment), Nonjudge (taking a nonevaluative stance to the internal environment), and Nonreact (letting go of thoughts and feelings). Later, Baer et al. ([<reflink idref="bib8" id="ref21">8</reflink>]) also validated FFMQ in meditating and nonmeditating samples. Previous research found that most mindfulness facets are significantly related to meditation training, various psychological symptoms, and the well‐being of a person (Baer et al., [<reflink idref="bib8" id="ref22">8</reflink>]; Fino et al., [<reflink idref="bib34" id="ref23">34</reflink>]; Goldberg et al., [<reflink idref="bib40" id="ref24">40</reflink>]; Roemer et al., [<reflink idref="bib79" id="ref25">79</reflink>]). As most of these studies used standard statistical techniques, Lecuona et al. ([<reflink idref="bib63" id="ref26">63</reflink>]) stressed the importance of using optimal statistical techniques to strengthen existing work on FFMQ.</p> <p>Given the apparent advantages of FFMQ, various short forms were also developed, such as the Short Inventory of Mindfulness Capability (12 items; Duan & Li, [<reflink idref="bib29" id="ref27">29</reflink>]) and the Balanced Inventory of Mindfulness‐related Skills (BIMS—27 items; Padmanabham et al., [<reflink idref="bib75" id="ref28">75</reflink>]). Baer et al. ([<reflink idref="bib4" id="ref29">4</reflink>]) also developed a short form—FFMQ (15 items)—to help reduce the participants' burden while adopting a unique construction method. The authors picked three items from each facet in their original research that described the domain well, had the highest factor loadings, and grouped them under respective domains. To the best of our knowledge, there is a dearth of research on the factor structure of FFMQ (15 items), and only three studies examined it (Feliu‐Soler et al., [<reflink idref="bib31" id="ref30">31</reflink>]; Gu et al., [<reflink idref="bib43" id="ref31">43</reflink>]; Kim et al., [<reflink idref="bib57" id="ref32">57</reflink>]), which led us to investigate it further in Pakistan using optimal statistical techniques.</p> <p>Despite numerous advantages, scholars reported a few issues against FFMQ. The foremost is the simplistic operationalization of the Buddhist concept—<emph>sati</emph> (Pali language)—through FFMQ without any theoretical basis while relying on standard statistical techniques (Bergomi et al., [<reflink idref="bib11" id="ref33">11</reflink>]). The empirical work in formulating FFMQ resulted in diminishing and altering the original Buddhist concept (Grossman, [<reflink idref="bib41" id="ref34">41</reflink>]), differential item functioning (Van Dam et al., [<reflink idref="bib91" id="ref35">91</reflink>]), and method effects (Van Dam et al., [<reflink idref="bib92" id="ref36">92</reflink>]). Though Baer et al. ([<reflink idref="bib5" id="ref37">5</reflink>]) found no signs of differential functioning in matched pairs of meditators and nonmeditators, Grossman and Van Dam ([<reflink idref="bib42" id="ref38">42</reflink>]) insisted that a different cut‐off point would exhibit differential functioning. As for method effect, Baer ([<reflink idref="bib3" id="ref39">3</reflink>]) defended using positive and negative statements as standard practice to avoid user acquiescence. In addition to issues related to FFMQ in general, the construction of FFMQ (15 items) is arguable because it did not adhere to the norms; for example, the overlap of long and short forms was not estimated (Smith et al., [<reflink idref="bib84" id="ref40">84</reflink>]). Moreover, the chosen items for FFMQ (15 items) might have inflated loadings in the original research due to the convergence process employed by computer software to find the best fit (Brown, [<reflink idref="bib16" id="ref41">16</reflink>]).</p> <p>FFMQ facets are also not without criticism. The Observe facet displayed poor psychometric properties among non‐meditators and positively correlated with maladaptive constructs, including dissociation, absent mindfulness, and thought suppression (Baer et al., [<reflink idref="bib7" id="ref42">7</reflink>]; Carpenter et al., [<reflink idref="bib22" id="ref43">22</reflink>]; Lilja et al., [<reflink idref="bib65" id="ref44">65</reflink>]; Tran et al., [<reflink idref="bib90" id="ref45">90</reflink>]). The altered behavior of the Observe facet might have occurred due to the absence of items related to emotional awareness (Rudkin et al., [<reflink idref="bib82" id="ref46">82</reflink>]), item wordings that do not reflect Buddhist practices (Feng et al., [<reflink idref="bib32" id="ref47">32</reflink>]), and the respondents' inability to pay unbiased attention (Baer et al., [<reflink idref="bib8" id="ref48">8</reflink>]). Besides Observe, the Describe facet was also problematic (Christopher et al., [<reflink idref="bib25" id="ref49">25</reflink>]; Feng et al., [<reflink idref="bib32" id="ref50">32</reflink>]; Grossman & Van Dam, [<reflink idref="bib42" id="ref51">42</reflink>]; Raman et al., [<reflink idref="bib78" id="ref52">78</reflink>]) and showed good psychometric properties in only 4.35% of the samples in a recent cross‐cultural study (Karl et al., [<reflink idref="bib56" id="ref53">56</reflink>]). Nonreact is yet another facet that showed a weak correlation with mindfulness in nonmeditators (Baer et al., [<reflink idref="bib7" id="ref54">7</reflink>]) to the extent that Tran et al. ([<reflink idref="bib90" id="ref55">90</reflink>]) suggested revising the content to improve item readability and discreteness.</p> <p>The higher‐order structure of FFMQ is also not without controversy. In contrast to the original single hierarchical model underpinned by five facets, scholars reported varying high‐order models, for example, the bifactor model (Aguado et al., [<reflink idref="bib1" id="ref56">1</reflink>]). Even in a single higher‐order model, Fernandez et al. ([<reflink idref="bib33" id="ref57">33</reflink>]) reported reversed directions for Actaware and Nonjudge facets. As for the existing three studies of FFMQ (15 items), the authors reported different models: (a) a five‐factor hierarchical structure with a single method factor (Kim et al., [<reflink idref="bib57" id="ref58">57</reflink>]), (b) a four‐factor hierarchical structure without Observe facet (Gu et al., [<reflink idref="bib43" id="ref59">43</reflink>]), and (c) a four‐factor bifactor model without Observe facet (Feliu‐Soler et al., [<reflink idref="bib31" id="ref60">31</reflink>]).</p> <p>In contrast to a single second‐order mindfulness factor, Tran et al. ([<reflink idref="bib90" id="ref61">90</reflink>]) proposed a two‐factor higher‐order model for nonmeditators based on a two‐part operational definition of mindfulness (Bishop et al., [<reflink idref="bib13" id="ref62">13</reflink>]). Self‐regulation (SR) refers to maintaining attention in the present moment; orientation to experience (OE) refers to accepting the present moment with curiosity and openness. The two‐factor higher‐order model provided a parsimonious theoretical explanation of the five‐facet structure and was further validated in the general population (Burzler et al., [<reflink idref="bib19" id="ref63">19</reflink>]). SR subsumes Observe and, among meditators, Nonreact and Describe; OE subsumes Actaware and Nonjudge (Burzler et al., [<reflink idref="bib19" id="ref64">19</reflink>]; Tran et al., [<reflink idref="bib90" id="ref65">90</reflink>]). The strength and direction of the correlation between SR and OE depend on the meditation experience of the sample. Unlike the original FFMQ, the two‐factor higher‐order FFMQ model has a theoretical basis and was intended for nonmeditators. Since this model was not tested before and earlier studies reported problems with the structure and showed various factor structures, we were interested in examining FFMQ (15 items) in Pakistan using optimal techniques.</p> <p>As self‐reported scales might alter their structure in other cultures (Henrich et al., [<reflink idref="bib48" id="ref66">48</reflink>]; Kalkbrenner, [<reflink idref="bib55" id="ref67">55</reflink>]), Karl et al. ([<reflink idref="bib56" id="ref68">56</reflink>]) highlighted that the current conceptualization of FFMQ suits western countries and stressed cross‐culture investigation of FFMQ. The most critical cultural dimension is the individualistic–collectivistic continuum (Hofstede, [<reflink idref="bib51" id="ref69">51</reflink>]; Lykes & Kemmelmeier, [<reflink idref="bib67" id="ref70">67</reflink>]). Individualism refers to an informal social network where people are more concerned about themselves. Whereas collectivism refers to a tightly knit social bond in which people are also concerned about extended families. Building on the individualism versus collectivism dimension, the self‐construal theory of Markus and Kitayama ([[<reflink idref="bib69" id="ref71">69</reflink>]]) theorized how culture conditioned people to believe and behave differently. In an independent view of self, people believe themselves to be independent and autonomous, with unique abilities and preferences that remain constant across contexts. People with an interdependent view prefer to fulfill obligations towards others; thus, their behaviors do not remain constant across contexts. Though the behavioral patterns do not remain constant (Gardner et al., [<reflink idref="bib36" id="ref72">36</reflink>]), culture impacts response motivation (King & McInerney, [<reflink idref="bib58" id="ref73">58</reflink>]). Therefore, western cultures mostly exhibit an independent view, and many non‐western cultures exhibit an interdependent view of self. Since Pakistan is a collectivist country (Country Comparison, [<reflink idref="bib26" id="ref74">26</reflink>]; Islam, [<reflink idref="bib53" id="ref75">53</reflink>]) and no cross‐culture examination of FFMQ (15 items) has yet been conducted, we also intended to add to the existing body of research while providing a cultural explanation of our results and comparing the data with the Spanish data (Feliu‐Soler et al., [<reflink idref="bib31" id="ref76">31</reflink>]).</p> <p>Several aspects necessitated and guided this study, adding to the existing body of research on FFMQ (15 items). First, despite the interest of scholars in studying mindfulness in the educational sector while using FFMQ, no study has yet investigated the structure of FFMQ (15 items) in collectivist Pakistan. Examining the factor structure in a different culture using optimal techniques is essential to prevent inadvertent measurement of mindfulness in applied settings, including the educational sector, especially once a wide range of theoretical, structural, statistical, and cultural issues are reported against FFMQ in general and FFMQ (15 items) in particular. Second, only three studies have reported using FFMQ (15 items), which warrants further studies. Third, the lack of cross‐cultural examination of FFMQ (15 items) hints a gap in the existing literature that asks for more studies about it. Based on these critical requirements, our work methodology catered to various aspects of the investigation and was as follows:</p> <p></p> <ulist> <item> 1. Evaluate the structure of FFMQ (15 items) in Pakistan through an EFA.</item> <p></p> <item> 2. Search the best fit model using CFA.</item> <p></p> <item> 3. Compare differences between males and females in mindfulness through multivariate analysis. Since the method effect was reported, estimate the impact of the polarity of items across gender through multigroup CFA (MGCFA).</item> <p></p> <item> 4. Conduct a cross‐cultural analysis of the current and Spanish data through multivariate analysis and MGCFA.</item> <p></p> <item> 5. Provide the cultural interpretation of the current data.</item> </ulist> <p>Following Feliu‐Soler et al. ([<reflink idref="bib31" id="ref77">31</reflink>]), we administered the FFMQ (15 items) directly to the sample.</p> <hd id="AN0164066401-3">METHOD</hd> <p></p> <hd id="AN0164066401-4">Data collection</hd> <p>The project had the approval of the Department of Psychology committee for research. Ten psychology undergrad students were trained in data collection, and they collected data from the students in 25 different departments in Pakistan's oldest public sector university during class timings from August to October 2021. The data collectors visited different classrooms during class timings and explained to the prospective students the purpose of the research and the method to complete various forms. A data dictionary was provided to help the participants' students understand different terms. We paid stipends and gave certificates of participation to the data collectors but did not pay the participants.</p> <hd id="AN0164066401-5">Participants</hd> <p>Five hundred eighty‐six nonmeditator and ethnically diverse students (<emph>M</emph><subs>age</subs> = 21.29 years, SD = 2.07) voluntarily completed the consent form and the FFMQ (15 items) using paper and pen. The students self‐identified themselves as Urdu speaking (53%), Punjabi (10%), Sindi (10%), and other ethnic minorities (27%). Data for the socioeconomic status was not collected. Eighty percent of the students reported intermediate level, whereas 20% showed advanced‐level proficiency in the English language. Approximately 81% of students were undergrad, less than 19% were in master classes, and less than 1% were postgraduate students.</p> <hd id="AN0164066401-6">Measure</hd> <p>Following Feliu‐Soler et al. ([<reflink idref="bib31" id="ref78">31</reflink>]), we administered the FFMQ (15 items) directly to avoid validation issues (Smith et al., [<reflink idref="bib84" id="ref79">84</reflink>]). FFMQ (15 items) contains five facets of mindfulness, including Observe (Items 1, 6, 11), Describe (Items 2, 7, 12), Actaware (Items 3, 8, 13), Nonjudge (Items 4, 9, 14), and Nonreact (Items 5, 10, 15). Participants responded using a 5‐point Likert scale (1 = never or very rarely true to 5 = very often or always true). The data were reverse‐scored appropriately before analysis. Previous literature has reported internal consistency measures (alpha) between.64 and.83 for each of the five facets of mindfulness.</p> <hd id="AN0164066401-7">Data analyses</hd> <p>We cleaned the data thoroughly using SPSS (V.23) (Osborne, [<reflink idref="bib74" id="ref80">74</reflink>]; Tabachnick & Fidell, [<reflink idref="bib87" id="ref81">87</reflink>]). Twenty‐one and nineteen cases were removed for more than 5% of missing data and unengaged responses. We employed an expectation–maximization procedure to impute the rest of the missing data. Since few variables were slightly non‐normal, various data transformation techniques were employed, including two‐step transformation (Templeton, [<reflink idref="bib89" id="ref82">89</reflink>]). As the data did not improve, we kept the data unchanged. The multivariate outlier analysis led to the removal of 9 more cases leaving behind 537 cases for final analysis (female: 407, 76%; male: 130, 24%).</p> <p>The initial CFA did not show acceptable fit indices, so we resorted to EFA. The sample was divided into two halves while keeping gender equality (Bandalos & Finney, [<reflink idref="bib9" id="ref83">9</reflink>]; Knekta et al., [<reflink idref="bib61" id="ref84">61</reflink>]). The EFA was conducted on polychoric data from the first half (<emph>n</emph> = 268: female = 203, 76%; male = 65, 24%) using the psych package in R (Revelle, [<reflink idref="bib81" id="ref85">81</reflink>]). Following Hair et al. ([<reflink idref="bib46" id="ref86">46</reflink>]), we removed items with low mean sampling adequacy value (MSA < 0.5) and low communality. The number of factors was determined using latent root criterion (eigenvalue > 1), scree test, and parallel test. As Leandre et al. ([<reflink idref="bib62" id="ref87">62</reflink>]) suggested for slightly non‐normal data, we employed principal axis factoring and direct oblimin rotational techniques. Items were also removed based on their subjective meaning in a particular factor. We calculated ordinal alpha (Gadermann et al., [<reflink idref="bib35" id="ref88">35</reflink>]) and omega (McNeish, [<reflink idref="bib71" id="ref89">71</reflink>]) for internal consistency. However, we also calculated mean inter‐item correlation (range:.2–.4) for internal consistency due to fewer indicators per facet (Pallant, [<reflink idref="bib76" id="ref90">76</reflink>]).</p> <p>The CFA was conducted on polychoric data from the second half (<emph>n</emph> = 269: female = 204, 76%; male = 65, 24%) using WLSMV estimator and the Lavaan package in R (Rosseel, [<reflink idref="bib80" id="ref91">80</reflink>]). We sequentially fitted five models, including (a) a one‐factor model where all the indicator variables were associated with one factor, (b) a correlated first‐order factor model with intercorrelation among factors, (c) a hierarchical model in which the first‐order factors were associated with a single second‐order mindfulness factor, (d) a bifactor model where the indicator variables were associated with their particular domain facet and were also associated with a single mindfulness factor, (e) a two‐factor model where first‐order factors were associated with two second‐order facets: SR and OE. We employed a two‐step methodology for hierarchical models (Brown, [<reflink idref="bib16" id="ref92">16</reflink>]). The first‐order correlated model was inspected for areas of strain in the first step; the hierarchical models were built on the first‐order correlated model in the second step. Consistent with the recommendations in the extant literature, the following multiple indices were used: a small near to zero <emph>χ</emph><sups>2</sups>; less than 2 [<emph>χ</emph><sups>2</sups>/<emph>df</emph>] ratio; comparative fit indices (CFI) > 0.95 as good and >0.90 as acceptable; Tucker–Lewis index (TLI) > 0.95 as good and >0.90 as acceptable; root mean square error of approximation (RMSEA) < 0.06 as good and <0.08 as acceptable; standardized root mean square residual (SRMR) < 0.08 as good and <0.10 as acceptable (Brown & Cudeck, [<reflink idref="bib15" id="ref93">15</reflink>]; Hair et al., [<reflink idref="bib46" id="ref94">46</reflink>]; Hu & Bentler, [<reflink idref="bib52" id="ref95">52</reflink>]). Heywood cases were adjusted by applying a low value (0.05). In addition, we examined standardized loadings (>0.3) and directionality to investigate the factor structure (Brown, [<reflink idref="bib16" id="ref96">16</reflink>]; Byrne, [<reflink idref="bib20" id="ref97">20</reflink>]). We computed the coefficient <emph>H</emph> (≥0.60) for reliability estimates to estimate the maximum reliability based on an optimally weighted scale (Feliu‐Soler et al., [<reflink idref="bib31" id="ref98">31</reflink>]; Kalkbrenner, [<reflink idref="bib55" id="ref99">55</reflink>]; McNeish, [<reflink idref="bib71" id="ref100">71</reflink>]).</p> <p>Using Hair et al. ([<reflink idref="bib46" id="ref101">46</reflink>]) four‐stage procedure, we conducted one‐way between group analysis (MANOVA) to estimate differences in mindfulness between males and females. Furthermore, we carried out MGCFA to evaluate the impact of the polarity of the items across gender. A new two‐factor correlated model was created using the R package Lavaan for ordinal variables (Hirschfeld & Von Brachel, [<reflink idref="bib49" id="ref102">49</reflink>]; Rosseel, [<reflink idref="bib80" id="ref103">80</reflink>]). The positive items corresponded to the first facet; the negative items corresponded to the second facet.</p> <p>Again, using Hair et al. ([<reflink idref="bib46" id="ref104">46</reflink>]) four‐stage procedure, we conducted MANOVA to estimate differences between Pakistan and the Spanish Data. In the second MGCFA, we estimated and compared correlated, one‐factor, hierarchical, bifactor, and two‐factor models between Pakistan and Spanish samples. A small CFI difference (≤0.01) indicates measurement invariance (Kline, [<reflink idref="bib60" id="ref105">60</reflink>]; Van de Schoot et al., [<reflink idref="bib93" id="ref106">93</reflink>]).</p> <p>Finally, we provided the cultural interpretation of the results.</p> <hd id="AN0164066401-8">RESULTS</hd> <p></p> <hd id="AN0164066401-9">Exploratory factor analysis</hd> <p>Following Hair et al. ([<reflink idref="bib46" id="ref107">46</reflink>]), two items of Describe facet (DES_2: <emph>I'm good at finding words to describe my feelings</emph>; DES_12: <emph>Even when I'm feeling terribly upset I can find a way to put it into words</emph>) and one item of Nonreact (NONR_15: When I have distressing thoughts or images I just notice them and let them go) were removed due to their low MSA and communality values, respectively. The latent root criterion, scree test, and the parallel test indicated four factors (Figure 1). Furthermore, we removed the last item of the Describe (DES_7R: <emph>I have trouble thinking of the right words to express how I feel about things</emph>) and 1 item of the Observe facet (OBS_6: <emph>I notice how foods and drinks affect my thoughts, bodily sensations, and emotions</emph>) that loaded wrongly on other factors. Finally, four factors (10 items) that describe 37% of the variance (Table 1) emerged. Mean inter‐item correlation was within the range (.2–.4), and ordinal alpha ranged from.43 to.61. The Observe and Nonreact facets were weakly positively correlated; Actaware and Nonjudge were moderately positively correlated (Table 2). Furthermore, while Nonjudge was weakly negatively correlated with both Observe and Nonreact, Actaware negatively correlated with Observe only. We retained this four‐facets (10 items) structure for further analysis.</p> <p> <img src="https://imageserver.ebscohost.com/img/embimages/rdk/PIS/01jul23/pits22875-fig-0001.jpg?ephost1=dGJyMNHX8kSepq84v%2bvlOLCmsE6epq5Srqa4SK6WxWXS" alt="pits22875-fig-0001.jpg" title="1 Determination of factors." /> </p> <p></p> <p>1 Table Exploratory factor analysis—Factor loadings for the 10 items of Four Facets of Mindfulness Questionnaire.</p> <p> <ephtml> <table><thead valign="bottom"><tr valign="bottom"><th align="left">Items</th><th align="left">NONJ</th><th align="left">ACT</th><th align="left">OBS</th><th align="left">NONR</th></tr></thead><tbody valign="top"><tr><td align="left">14R. I tell myself I shouldn't be feeling the way I'm feeling</td><td align="char" char=".">0.67</td><td /><td /><td /></tr><tr><td align="left">4R. I believe some of my thoughts are abnormal or bad and I shouldn't think that way</td><td align="char" char=".">0.56</td><td /><td /><td /></tr><tr><td align="left">9R. I think some of my emotions are bad or inappropriate and I shouldn't feel them</td><td align="char" char=".">0.42</td><td /><td /><td /></tr><tr><td align="left">8R. I do jobs or tasks automatically without being aware of what I'm doing</td><td /><td align="char" char=".">0.69</td><td /><td /></tr><tr><td align="left">3R. I don't pay attention to what I'm doing because I'm daydreaming, worrying, or otherwise distracted</td><td /><td align="char" char=".">0.50</td><td /><td /></tr><tr><td align="left">13R. I find myself doing things without paying attention</td><td /><td align="char" char=".">0.40</td><td /><td /></tr><tr><td align="left">11. I pay attention to sensations, such as the wind in my hair or sun on my face</td><td /><td /><td align="char" char=".">0.71</td><td /></tr><tr><td align="left">1. When I take a shower or a bath, I stay alert to the sensations of water on my body</td><td /><td /><td align="char" char=".">0.55</td><td /></tr><tr><td align="left">10. When I have distressing thoughts or images I am able just to notice them without reacting.</td><td /><td /><td /><td align="char" char=".">0.44</td></tr><tr><td align="left">5. When I have distressing thoughts or images, I "step back" and am aware of the thought or image without getting taken over by it</td><td /><td /><td /><td align="char" char=".">0.35</td></tr><tr><td align="left">SS Loadings</td><td align="char" char=".">1.2</td><td align="char" char=".">1.07</td><td align="char" char=".">1.02</td><td align="char" char=".">0.44</td></tr><tr><td align="left">% Cumulative factor variance</td><td align="char" char=".">0.12</td><td align="char" char=".">0.23</td><td align="char" char=".">0.33</td><td align="char" char=".">0.37</td></tr><tr><td align="left">Omega</td><td align="char" char=".">.60</td><td align="char" char=".">.60</td><td align="char" char=".">.55</td><td align="char" char=".">.43</td></tr><tr><td align="left">Ordinal α</td><td align="char" char=".">.61</td><td align="char" char=".">.59</td><td align="char" char=".">.55</td><td align="char" char=".">.43</td></tr><tr><td align="left">Mean inter‐item correlation</td><td align="char" char=".">.30</td><td align="char" char=".">.28</td><td align="char" char=".">.34</td><td align="char" char=".">.24</td></tr></tbody></table> </ephtml> </p> <p>1 <emph>Note</emph>: NONJ: Nonjudge, Items 14R, 4R, 9R; ACT; Actaware, Items 3R, 8R, 13R; OBS: Observe, Items 1, 11; NONR: Nonreact, Items 5, 10; R denotes reverse coded; polychoric correlation was used to calculate Ordinal <emph>α</emph>.</p> <p>2 Table Summary of mean (SD) and Intercorrelation between facets after EFA.</p> <p> <ephtml> <table><thead valign="bottom"><tr valign="bottom"><th /><th align="left">Observe</th><th align="left">Nonreact</th><th align="left">Actaware</th><th align="left">Nonjudge</th></tr></thead><tbody valign="top"><tr><td align="left">Observe</td><td align="left">1.00</td><td align="left">0.19</td><td align="char" char=".">−0.23</td><td align="char" char=".">−0.29</td></tr><tr><td align="left">Nonreact</td><td /><td align="left">1.00</td><td align="char" char=".">0.14</td><td align="char" char=".">−0.26</td></tr><tr><td align="left">Actaware</td><td /><td /><td align="char" char=".">1.00</td><td align="char" char=".">0.42</td></tr><tr><td align="left">Nonjudge</td><td /><td /><td /><td align="char" char=".">1.00</td></tr><tr><td align="left">Mean (SD)</td><td align="left">3.25(1.04)</td><td align="left">3.01 (0.86)</td><td align="char" char=".">3.34 (0.933)</td><td align="char" char=".">3.01 (0.91)</td></tr></tbody></table> </ephtml> </p> <p>2 Abbreviation: EFA, exploratory factor analysis.</p> <hd id="AN0164066401-11">Confirmatory factor analysis</hd> <p>A two‐step CFA strategy for hierarchical models was employed (Brown, [<reflink idref="bib16" id="ref108">16</reflink>]). In the first step, the correlated model provided good fit statistics (<emph>χ</emph><sups>2</sups> (df) = 46.36 (<reflink idref="bib29" id="ref109">29</reflink>), CFI = 0.981, TLI = 0.970, RMSEA = 0.039 [0.015–0.060], SRMR = 0.044; Table 3). Moreover, the standardized factor loadings were greater than 0.3 (range: 0.439–0.644), and the directionality was as expected. Observe and Nonreact moderately correlated with each other (0.458) and negatively correlated with Actaware and Nonjudge. On the other hand, Actaware correlated positively and strongly with the Nonjudge facet (0.710). The specific result hinted that the two‐factor superordinate structure existed in the data (Burzler et al., [<reflink idref="bib19" id="ref110">19</reflink>]; Tran et al., [<reflink idref="bib90" id="ref111">90</reflink>]), comprising positively worded facets (Observe, Nonreact) and negatively worded facets (Actaware, Nonjudge).</p> <p>3 Table Fit Indices of the five models of FFMQ (n = 269) using polychoric data.</p> <p> <ephtml> <table><thead valign="bottom"><tr valign="bottom"><th /><th align="left"><italic>χ</italic><sup>2</sup></th><th align="left">df</th><th align="left"><italic>χ</italic><sup>2</sup>/df</th><th align="left">CFI</th><th align="left">TLI</th><th align="left">RMSEA [90%]</th><th align="left">SRMR</th></tr></thead><tbody valign="top"><tr><td align="left">Correlated four factor</td><td align="char" char=".">32.19</td><td align="char" char=".">29</td><td align="char" char=".">1.11</td><td align="char" char=".">0.995</td><td align="char" char=".">0.992</td><td align="left">0.020 [0.00–0.052]</td><td align="char" char=".">0.044</td></tr><tr><td align="left">One factor</td><td align="char" char=".">83.05</td><td align="char" char=".">35</td><td align="char" char=".">2.37</td><td align="char" char=".">0.923</td><td align="char" char=".">0.901</td><td align="left">0.072 [0.052–0.092]</td><td align="char" char=".">0.070</td></tr><tr><td align="left">Hierarchical</td><td align="char" char=".">37.19</td><td align="char" char=".">32</td><td align="char" char=".">1.16</td><td align="char" char=".">0.992</td><td align="char" char=".">0.988</td><td align="left">0.025 [0.00–0.054]</td><td align="char" char=".">0.048</td></tr><tr><td align="left">Bifactor</td><td align="char" char=".">24.62</td><td align="char" char=".">27</td><td align="char" char=".">.912</td><td align="char" char=".">1.00</td><td align="char" char=".">1.00</td><td align="left">0.00 [0.00–0.042]</td><td align="char" char=".">0.040</td></tr><tr><td align="left">Two‐factor hierarchical</td><td align="char" char=".">34.79</td><td align="char" char=".">31</td><td align="char" char=".">1.12</td><td align="char" char=".">0.994</td><td align="char" char=".">0.991</td><td align="left">0.021 [0.00–0.052]</td><td align="char" char=".">0.046</td></tr></tbody></table> </ephtml> </p> <ulist> <item>3 <emph>Note</emph> : Indices with italics fall within acceptable range.</item> <item>4 Abbreviations: <emph>χ</emph><sups>2</sups>, Chi square; CFI, comparative fit indices; CI, confidence interval; FFMQ, Five Facets Mindfulness Questionnaire; RMSEA, root mean square error of approximation; SRMR, standardized root mean square residual; TLI, Tucker–Lewis index.</item> </ulist> <p>In the second step of CFA, we tested four different models: one‐factor, hierarchical, bifactor, and two‐factor hierarchical models. The one‐factor, hierarchical, and bifactor models provided good fit indices (Table 3) but failed on other criteria: factor loadings were less than 0.3, and a few indicators loaded negatively. Finally, we tested two‐factor hierarchical model (Figure 2), which produced good fit results (<emph>χ</emph><sups>2</sups> (df) = 34.79 (<reflink idref="bib31" id="ref112">31</reflink>), CFI = 0.994, TLI = 0.991, RMSEA = 0.021 [0.00–0.052], SRMR = 0.046; Table 3) and satisfied other criteria: factor loadings (>0.3) and directionality was as per expectation. The reliability analysis showed a high coefficient <emph>H</emph> of both higher‐order factors (SA = 0.84; OE = 0.89). The negative correlation between the two higher‐order factors (<emph>r</emph> = −.79) indicated that these two factors act independently in the current nonmeditator sample.</p> <p> <img src="https://imageserver.ebscohost.com/img/embimages/rdk/PIS/01jul23/pits22875-fig-0002.jpg?ephost1=dGJyMNHX8kSepq84v%2bvlOLCmsE6epq5Srqa4SK6WxWXS" alt="pits22875-fig-0002.jpg" title="2 Two‐factors hierarchical model." /> </p> <p></p> <hd id="AN0164066401-13">One‐way MANOVA between males and females</hd> <p>Since positive correlations between various facets resulted in two separate groups (Observe and Nonreact; Actaware and Nonjudge), we conducted two MANOVAs. In the first MANOVA, we investigated the gender differences in Observe and Nonreact facets; in the second MANOVA, we investigated the gender differences in Actaware and Nonjudge facets. Preliminary investigations did not result in any serious violations. In the first MANOVA, there was a statistically significant difference between males and females on the combined dependent variables (Observe and Nonreact), <emph>F</emph>(<reflink idref="bib2" id="ref113">2</reflink>, 266) = 4.79, <emph>p</emph> = .009; Wilk's Lambda = 0.965; partial eta squared = 0.035. We investigated the univariate analysis using Bonferroni adjusted alpha level of.025, and only Nonreact reached statistical significance, <emph>F</emph>(<reflink idref="bib1" id="ref114">1</reflink>, 268) = 5.643, <emph>p</emph> = .018; partial eta squared = 0.021. The mean scores of Nonreact indicated that males reported higher level (mean = 6.47, SD = 0.231) than females (mean = 5.84, SD = 0.130). In the second MANOVA, we did not find any statistically significant difference between males and females on the combined variables (Actaware and Nonjudge), <emph>F</emph>(<reflink idref="bib2" id="ref115">2</reflink>, 266) = 0.27, <emph>p</emph> = .759; Wilk's Lambda = 0.8; partial eta squared = 0.002. Consequently, the univariate analysis did not yield any significant differences between males and females on Actaware and Nonjudge.</p> <hd id="AN0164066401-14">MGCFA to investigate impact of polarity of items across gender</hd> <p>The MGCFA result to estimate the impact of the polarity of items across gender indicates that only metric invariance is achieved (Table 4). The adjusted scalar invariance could not produce the desired ΔCFI (<0.01) cut‐off value; therefore, strict invariance was not measured.</p> <p>4 Table Multigroup CFA for the polarity of items across gender.</p> <p> <ephtml> <table><thead valign="bottom"><tr valign="bottom"><th /><th align="left"><italic>χ</italic><sup>2</sup></th><th align="left">df</th><th align="left"><italic>χ</italic><sup>2</sup>/df</th><th align="left">CFI</th><th align="left">ΔCFI</th><th align="left">TLI</th><th align="left">RMSEA [90%]</th><th align="left">SRMR</th></tr></thead><tbody valign="top"><tr><td align="left">Configural</td><td align="char" char=".">109.32</td><td align="char" char=".">68</td><td align="char" char=".">1.60</td><td align="char" char=".">0.942</td><td align="left">–</td><td align="char" char=".">0.924</td><td align="left">0.067 [0.043, 0.09]</td><td align="char" char=".">0.06</td></tr><tr><td align="left">Metric</td><td align="char" char=".">117.50</td><td align="char" char=".">76</td><td align="char" char=".">1.54</td><td align="char" char=".">0.942</td><td align="left">0</td><td align="char" char=".">0.931</td><td align="left">0.064 [0.040, 0.08]</td><td align="char" char=".">0.08</td></tr><tr><td align="left">Scalar0002</td><td align="char" char=".">160.18</td><td align="char" char=".">104</td><td align="char" char=".">1.54</td><td align="char" char=".">0.921</td><td align="left">0.021</td><td align="char" char=".">0.932</td><td align="left">0.064 [0.043, 0.08]</td><td align="char" char=".">0.08</td></tr></tbody></table> </ephtml> </p> <ulist> <item>5 Abbreviations: ACT, Actaware, reverse coded; NONJ, Nonjudge, reverse coded; NONR, Nonreact; OBS, Observe; OE, Orient to Experience (Actaware, Nonjudge); SR, Self‐Regulation (Observer, Nonreact).</item> <item>6 a Adjusted scalar invariance.</item> </ulist> <hd id="AN0164066401-15">One‐way MANOVA between Pakistan and Spanish data</hd> <p>Again, we conducted two MANOVAs as a positive correlation existed in our data between various facets (Observe and Nonreact; Actaware and Nonjudge). Preliminary investigations did not result in any serious violations. In the first MANOVA, we observed there was no statistically significant difference between samples from Pakistan and Spanish, <emph>F</emph>(<reflink idref="bib2" id="ref116">2</reflink>, 1066) = 0.929, <emph>p</emph> = .395; Wilk's Lambda = 0.998; partial eta squared = 0.002. Consequently, there were no significant univariate differences. However, we observed in the second MANOVA that there was a statistically significant difference between Pakistan and Spanish data on the combined dependent variables (Actaware and Nonjudge), <emph>F</emph>(<reflink idref="bib2" id="ref117">2</reflink>, 1066) = 2.88, <emph>p</emph> < .001; Wilk's Lambda = 0.852, partial eta squared = 0.148. The univariate analysis, using Bonferroni adjusted alpha level of.025, indicated that Nonjudge reached statistical significance, <emph>F</emph>(<reflink idref="bib1" id="ref118">1</reflink>, 1067) = 149.138, <emph>p</emph> < .001; partial eta squared = 0.123. The mean scores indicated that Spaniards reported higher level of Nonjudge (mean = 11.45, SD = 0.102) than Pakistanis (mean = 9.00, SD = 0.176).</p> <hd id="AN0164066401-16">MGCFA to investigate fit of data between Pakistan and Spanish samples</hd> <p>Table 5 displays the results of the second MGCFA across Pakistan and Spanish samples. The two samples display configural invariance across most models, but the correlated four‐factors model also displayed metric invariance (ΔCFI < 0.01).</p> <p>5 Table Multigroup CFA (Spanish sample, n = 800; Pakistani sample, n = 269).</p> <p> <ephtml> <table><thead valign="bottom"><tr valign="bottom"><th /><th><italic>χ</italic><sup>2</sup></th><th>df</th><th><italic>χ</italic><sup>2</sup>/df</th><th>CFI</th><th>ΔCFI</th><th>TLI</th><th>RMSEA [90%]</th><th>SRMR</th></tr></thead><tbody valign="top"><tr><td align="left">One‐factor model</td><td /><td /><td /><td /><td /></tr><tr><td align="left">Configural</td><td align="char" char=".">147.57</td><td align="char" char=".">60</td><td align="char" char=".">2.45</td><td align="char" char=".">0.991</td><td align="left">–</td><td align="char" char=".">0.987</td><td align="left">0.052 [0.042–0.063]</td><td align="char" char=".">0.048</td></tr><tr><td align="left">Metric</td><td align="char" char=".">488.78</td><td align="char" char=".">70</td><td align="char" char=".">6.98</td><td align="char" char=".">0.959</td><td align="left">0.03</td><td align="char" char=".">0.947</td><td align="left">0.106 [0.097–0.115]</td><td align="char" char=".">0.079</td></tr><tr><td align="left">Correlated four factor</td><td /><td /><td /><td /><td /></tr><tr><td align="left">Configural</td><td align="char" char=".">330.42</td><td align="char" char=".">62</td><td align="char" char=".">5.32</td><td align="char" char=".">0.973</td><td align="left">–</td><td align="char" char=".">0.961</td><td align="left">0.090 [0.081–0.100]</td><td align="char" char=".">0.068</td></tr><tr><td align="left">Metric</td><td align="char" char=".">338.04</td><td align="char" char=".">66</td><td align="char" char=".">5.12</td><td align="char" char=".">0.973</td><td align="left">0</td><td align="char" char=".">0.961</td><td align="left">0.088 [0.079–0.097]</td><td align="char" char=".">0.069</td></tr><tr><td align="left">Scalar0003</td><td align="char" char=".">557.23</td><td align="char" char=".">92</td><td align="char" char=".">6.43</td><td align="char" char=".">0.954</td><td align="left">0.02</td><td align="char" char=".">0.955</td><td align="left">0.097 [0.090–0.105]</td><td align="char" char=".">0.070</td></tr><tr><td align="left">Hierarchical</td><td /><td /><td /><td /></tr><tr><td align="left">Configural</td><td align="char" char=".">437.78</td><td align="char" char=".">68</td><td align="char" char=".">6.43</td><td align="char" char=".">0.963</td><td align="left">–</td><td align="char" char=".">0.952</td><td align="left">0.101 [0.092–0.110]</td><td align="char" char=".">0.080</td></tr><tr><td align="left">Metric</td><td align="char" char=".">695.26</td><td align="char" char=".">75</td><td align="char" char=".">9.27</td><td align="char" char=".">0.939</td><td align="left">0.025</td><td align="char" char=".">0.926</td><td align="left">0.125 [0.126–0.133]</td><td align="char" char=".">0.100</td></tr><tr><td align="left">Bifactor</td><td /><td /><td /><td /><td /><td /><td /><td /></tr><tr><td align="left">Configural</td><td align="char" char=".">186.26</td><td align="char" char=".">56</td><td align="char" char=".">3.32</td><td align="char" char=".">0.987</td><td align="left">–</td><td align="char" char=".">0.979</td><td align="left">0.066 [0.056–0.077]</td><td align="char" char=".">0.052</td></tr><tr><td align="left">Metric</td><td align="char" char=".">367.21</td><td align="char" char=".">59</td><td align="char" char=".">6.22</td><td align="char" char=".">0.970</td><td align="left">0.017</td><td align="char" char=".">0.961</td><td align="left">0.090 [0.081–0.099]</td><td align="char" char=".">0.070</td></tr><tr><td align="left">Two‐factor hierarchical</td><td /><td /><td /><td /></tr><tr><td align="left">Configural</td><td align="char" char=".">386.22</td><td align="char" char=".">68</td><td align="char" char=".">5.67</td><td align="char" char=".">0.968</td><td align="left">–</td><td align="char" char=".">0.958</td><td align="left">0.090 [0.085–0.103]</td><td align="char" char=".">0.074</td></tr><tr><td align="left">Metric</td><td align="char" char=".">606.11</td><td align="char" char=".">76</td><td align="char" char=".">7.97</td><td align="char" char=".">0.947</td><td align="left">0.021</td><td align="char" char=".">0.938</td><td align="left">0.114 [0.106–0.123]</td><td align="char" char=".">0.093</td></tr></tbody></table> </ephtml> </p> <ulist> <item>7 <emph>Note</emph> : ΔCFI cut‐off is <0.01.</item> <item>8 Abbreviations: <emph>χ</emph><sups>2</sups> , Chi square; CFI, comparative fit indices; RMSEA, root mean square error of approximation; SRMR, standardized root mean square residual; TLI, Tucker–Lewis index.</item> <item>9 a Adjusted scalar invariance.</item> </ulist> <hd id="AN0164066401-17">DISCUSSION</hd> <p>This first systematic investigation of the least researched FFMQ (15 items) examined its factor structure using optimal techniques among nonmeditators' university students in Pakistan. In addition, it aimed to add to the existing body of research while providing a cultural explanation of our data and conducting the cross‐cultural examination of data between Pakistan and Spanish samples. FFMQ is an empirically derived scale, and a change in its structure is not unlikely (Baer, [<reflink idref="bib3" id="ref119">3</reflink>]). As for FFMQ (15 items), the studies on it are still in their infancy. To the author's knowledge, only three studies examined FFMQ (15 items) (Feliu‐Soler et al., [<reflink idref="bib31" id="ref120">31</reflink>]; Gu et al., [<reflink idref="bib43" id="ref121">43</reflink>]; Kim et al., [<reflink idref="bib57" id="ref122">57</reflink>]), and which were restricted to examining various models. In contrast, we adopted a more robust approach while conducting EFA and CFA using polychoric correlations. However, our work is identical to the others in one aspect; it revealed yet another structure of FFMQ (15 items). The EFA deleted Describe facet and revealed four facets (10 items): Observe, Nonreact, Actaware, and Nonjudge, showing good mean inter‐item correlation (range:.24–.30) as an internal consistency reliability measure. Based on the correlated first‐order factor model, only the two‐factor hierarchical model (SR: Observe and Nonreact; OE: Actaware and Nonjudge) produced good fit indices, adequate factor loadings, and maintained the directions. Both factors exhibited a strong reliability coefficient (coefficient <emph>H</emph> > 0.8) and were unrelated (<emph>r</emph> = −.79). The first MANOVA exhibited a nonsignificant difference between males and females in all facets of mindfulness, except for the Nonreact facet; the first MGCFA to determine the impact of the polarity of items across gender achieved metric invariance. In cross‐cultural analysis, the second MANOVA displayed a nonsignificant difference between the Pakistan and Spanish samples, except for the Nonjudge facet. The second MGCFA demonstrated configural invariance in most models and achieved metric invariance in the correlated four‐factor model. Several unique aspects of this study that add to the previous literature and cultural explanations of the results are discussed below.</p> <p>In line with Kim et al. ([<reflink idref="bib57" id="ref123">57</reflink>]), the current study retained the Observe facet. Though the Observe facet is a critical aspect of all definitions of mindfulness, prior research showed the inapplicability of the Observe facet in nonmeditators (Baer et al., [[<reflink idref="bib7" id="ref124">7</reflink>]]; Lilja et al., [<reflink idref="bib65" id="ref125">65</reflink>]). The general perspective of the Observe facet holds that meditation practices enhance and improve the nonjudgmental observing ability of an individual. Since nonmeditators do not indulge in formal meditation practices, the Observe facet does not appear in most studies. Two alternative explanations in favor of the Observe facet can be given. First, as Haas and Akamatsu ([<reflink idref="bib45" id="ref126">45</reflink>]) highlighted, culture might enhance the observing ability of a person. Therefore, it is likely that individuals in collectivist Pakistani society are more observant than others. We will deliberate on the cultural aspect later. Second, the theoretical aspect of the Observe facet ensures its inclusion in the mindfulness questionnaire. The noninclusion of the Observe facet diminishes the Buddhist concept of mindfulness, which is succinctly defined as <emph>Conscious awareness</emph> (Hanh, [<reflink idref="bib47" id="ref127">47</reflink>]) and <emph>lucid awareness</emph> (Bodhi, [<reflink idref="bib14" id="ref128">14</reflink>]). Both these terms tacitly highlight the importance of the Observe facet in FFMQ. Based on this reasoning, we believe that removing the Observe facet from FFMQ will alter the fundamental nature of FFMQ as a mindfulness questionnaire. Thus, similar to various earlier research, our result removed the current incongruity regarding Observe facet in FFMQ (Burzler et al., [<reflink idref="bib19" id="ref129">19</reflink>]; Kim et al., [<reflink idref="bib57" id="ref130">57</reflink>]; Tran et al., [<reflink idref="bib90" id="ref131">90</reflink>]).</p> <p>Another unique finding of the current research is the removal of the Describe facet. The Buddhist term <emph>sati</emph> (Pali language) was translated to mindfulness by T. W. Rhys Davids, and this translation is questionable since more appropriate translations, such as memory, remembrance, or recollections, were available (Bodhi, [<reflink idref="bib14" id="ref132">14</reflink>]; Gethin, [[<reflink idref="bib37" id="ref133">37</reflink>]]). The alternative translation suggests that pure awareness, remembering, and recollecting events without describing the internal or external experience reflect a Buddhist sense of mindfulness (Christopher et al., [<reflink idref="bib25" id="ref134">25</reflink>]; Feng et al., [<reflink idref="bib32" id="ref135">32</reflink>]). In other words, sustained mindfulness practices do not require describing the experience (Goldstein, [<reflink idref="bib39" id="ref136">39</reflink>]). However, it is one of the techniques, like breath counting (Levinson et al., [<reflink idref="bib64" id="ref137">64</reflink>]), to contact the experience at a preverbal level (Grossman & Van Dam, [<reflink idref="bib42" id="ref138">42</reflink>]), which might help a novice maintain contact for a more extended period. Moreover, the Describe facet was part of KIMS—a constituent scale for FFMQ—developed to assist in dialectical behavior therapy (DBT) (Linehan, [<reflink idref="bib66" id="ref139">66</reflink>]). Therefore, we believe that the Describe sub facet is theoretically limited to DBT only, and its usage in applied settings will alter the results. We describe the cultural explanation of the removal of the Describe facet later.</p> <p>Nonreact is another facet that requires consideration. Nonreact is a unique quality of mindfulness that requires experiencing an event without a reaction. However, as the definition suggests, it is unlikely for most nonmeditators to let go of an event without expressing a reaction. Earlier, Baer et al. ([<reflink idref="bib7" id="ref140">7</reflink>]) found a weak correlation of Nonreact with mindfulness. Van Dam et al. ([<reflink idref="bib91" id="ref141">91</reflink>]) also pointed out differential items functioning between meditators and nonmeditators in Nonreact. Even Kim et al. ([<reflink idref="bib57" id="ref142">57</reflink>]) reported low factor loadings (<0.4) of one of the items in Nonreact. In line with Tran et al. ([<reflink idref="bib90" id="ref143">90</reflink>]), we surmise that one of the reasons for the poor functioning of the Nonreact facet in nonmeditators is the items' incomprehensibility by the nonmeditators, which requires revision.</p> <p>The current study has observed various negative associations between the facets, leading to the two‐factors higher‐order model (SR: Observe and Nonreact; OE: Actaware and Nonjudge). A negative correlation is typical of nonmeditators, as positive correlations between the factors occur in the meditators (Baer et al., [[<reflink idref="bib7" id="ref144">7</reflink>]]). Unlike meditators, a negative association between Observe and Nonjudge facets shows the inclination of the sample to evaluate all experiences and their inability to maintain a nonjudgmental stance which hinders the development of mindfulness skills in them (Baer et al., [<reflink idref="bib6" id="ref145">6</reflink>]; Lilja et al., [<reflink idref="bib65" id="ref146">65</reflink>]; Watson‐Singleton et al., [<reflink idref="bib97" id="ref147">97</reflink>]). Similarly, a negative association between Observe and Actaware signifies how non‐meditators maladaptively apply attentional resources, identical to those with anxiety symptoms (Curtiss & Klemanski, [<reflink idref="bib28" id="ref148">28</reflink>]). Similar to the collectivist Bhutanese sample (Haas & Akamatsu, [<reflink idref="bib45" id="ref149">45</reflink>]), Nonreact and Nonjudge also correlated negatively in the current sample, which means a reactive tendency of the current nonmeditator sample while simultaneously evaluating events.</p> <p>Consistent with the prior research (Burzler et al., [<reflink idref="bib19" id="ref150">19</reflink>]; Tran et al., [<reflink idref="bib90" id="ref151">90</reflink>]), our data showed that the two‐factor higher‐order model best suits the nonmeditators. Moreover, a strong negative correlation between the two‐higher order structure—SR and OE—and the high reliability of each (coefficient <emph>H</emph> > 0.8) indicate that the two factors operate independently, and the total score cannot be computed. Since meditation training induces a nonreactive and nonevaluative stance (Burzler et al., [<reflink idref="bib19" id="ref152">19</reflink>]; Tran et al., [<reflink idref="bib90" id="ref153">90</reflink>]), we could observe a strong correlation between the higher‐order factors in a meditation sample.</p> <p>Another unique finding of the current study is that except for the Nonjudge facet, there is a nonsignificant difference in the means of the three latent variables (Observe, Nonreact, and Actaware) between Pakistan and the Spanish sample. Consistent with the earlier cross‐cultural research for Nonjudge (Christopher et al., [<reflink idref="bib24" id="ref154">24</reflink>]; Haas & Akamatsu, [<reflink idref="bib45" id="ref155">45</reflink>]; Sugiura et al., [<reflink idref="bib86" id="ref156">86</reflink>]), the current sample scored lesser than the Spanish sample. Nonjudge is about the nonevaluation of the experience, and a low score signifies the current sample's tendency to evaluate the circumstances. The evaluation of an event is the collectivist cultural exhibit of the notion of self‐criticism (Kitayama et al., [<reflink idref="bib59" id="ref157">59</reflink>]), which we discuss later. Congruent with the work of Haas and Akamatsu ([<reflink idref="bib45" id="ref158">45</reflink>]), we found configural invariance in all the models between Pakistan and Spanish samples, but metric invariance is achieved in a correlated four‐factor model only. One probable reason for the similarity in structure in the present study is that Spain is more collectivist than the rest of the European countries (Country Comparison, [<reflink idref="bib27" id="ref159">27</reflink>]). However, like other western countries, the Spaniards seem more individualistic in their personal lives. As Karl et al. ([<reflink idref="bib56" id="ref160">56</reflink>]) pointed out, our results further suggest that the current items in FFMQ appear more individualistic.</p> <p>The current results should be seen from a cultural angle as well. The self‐construal theory states that culture might condition people such that people in collectivist cultures exhibit tight social bonds and prefer to maintain strong and harmonious relationships with significant others (Markus & Kitayama, [<reflink idref="bib69" id="ref161">69</reflink>], [<reflink idref="bib70" id="ref162">70</reflink>]). In Pakistan, parents occupy a significant position (Zahra & Saleem, [<reflink idref="bib99" id="ref163">99</reflink>]), and children must show conformity, obedience, loyalty, and respect to them and accept their decisions unquestionably (Bibi et al., [<reflink idref="bib12" id="ref164">12</reflink>]; Stewart et al., [<reflink idref="bib85" id="ref165">85</reflink>]; Yasinzai et al., [<reflink idref="bib98" id="ref166">98</reflink>]; Zaman, [<reflink idref="bib101" id="ref167">101</reflink>]). Since culture changes response considerations (King & McInerney, [<reflink idref="bib58" id="ref168">58</reflink>]), we believe in the argument presented by Haas and Akamatsu ([<reflink idref="bib45" id="ref169">45</reflink>]) that the children in a collectivist society of Pakistan might have improved observing ability, which forces them to constantly observe their behavior to realign as per the wishes of the significant people: parents. Another critical cultural exhibit is that any independent behavior, for example, describing one's feelings, is seen as nonconformist and disrespectful (Markus & Kitayama, [<reflink idref="bib69" id="ref170">69</reflink>]; p. 227), especially if it is against the parent's wishes and which might jeopardize the delicate harmony in a family. Such displays can be seen culturally in many forms. For example, in a famous movie, <emph>three idiots</emph> (Hirani, [<reflink idref="bib50" id="ref171">50</reflink>]), one of the actors was reluctant to speak against his parent's decision about choosing a profession of his own choice. Therefore, we observed low congruence among respondents for the Describe facet leading to its extinction in the data.</p> <p>Similarly, the negative association between Nonreact and Nonjudge is a common collectivist cultural representation (Haas & Akamatsu, [<reflink idref="bib45" id="ref172">45</reflink>]; Menon et al., [<reflink idref="bib72" id="ref173">72</reflink>]; Virlley & Reddy, [<reflink idref="bib94" id="ref174">94</reflink>]). Whereas the individualist society nourishes self‐enhancement, the collectivist society promotes self‐criticism (Kitayama et al., [<reflink idref="bib59" id="ref175">59</reflink>]). Self‐criticism promotes unworthiness, failure, and guilt and leads self‐critical individuals to brutally evaluate themselves to seek acceptance from others and avoid being disapproved (Warren et al., [<reflink idref="bib96" id="ref176">96</reflink>]). In the collectivist society of Pakistan, self‐criticism serves as a primer to prevent shame and loss of face in front of parents (Stewart et al., [<reflink idref="bib85" id="ref177">85</reflink>]; Zaman, [<reflink idref="bib101" id="ref178">101</reflink>]). It is further deepened by the informal training or <emph>tarbiyat</emph> (moral code deepened by custom) and adherence to the religious traditions that constantly remind children of the importance of parents in a family and how they must act to maintain harmonious relationships with parents while restraining any overt act of dissent (Yasinzai et al., [<reflink idref="bib98" id="ref179">98</reflink>]; Zaman, [<reflink idref="bib101" id="ref180">101</reflink>]). In Pakistan, self‐criticism affects females more than males, and they exhibit more depressive symptoms than males (Tariq & Yousaf, [<reflink idref="bib88" id="ref181">88</reflink>]). Our results correspond to this finding once we found that females are less likely to let go of any bad experience and perhaps indulge more in criticizing themselves. Overall, our result is reminiscent of the cultural aspect of Pakistan's society.</p> <hd id="AN0164066401-18">IMPLICATIONS FOR RESEARCH AND PRACTICE</hd> <p>This study has implications for research and practice. FFMQ (39 items) or FFMQ (15 items) are empirically derived scales; therefore, a change in their structure is expected (Baer, [<reflink idref="bib3" id="ref182">3</reflink>]), primarily if it is being investigated in other cultures (Heinrich et al., 2010). This study found that FFMQ can be grounded in mindfulness theory (Bishop et al., [<reflink idref="bib13" id="ref183">13</reflink>]), and several authors have provided empirical support for it (Burzler et al., [<reflink idref="bib19" id="ref184">19</reflink>]; Tran et al., [<reflink idref="bib90" id="ref185">90</reflink>]). Moreover, the study has provided sufficient reasons for excluding Describe facet in a collectivist society. Developing culturally accepted items for the Observe and Nonreact might improve their reliabilities.</p> <p>Based on this first investigative study, it would be early to make any strong suggestion regarding the potential use of FFMQ (15 items). As the impact of mindfulness on students has been researched extensively in schools, school psychologists might further test the structure of FFMQ (15 items) in other student populations. However, it must be noted that the meditation experience would change the relationship of the higher‐order structure (SR and OE). Instead of computing the total score, the nonmeditators might still be scored separately on two higher‐order dimensions.</p> <hd id="AN0164066401-19">CONCLUSION, LIMITATION, AND FUTURE DIRECTIONS</hd> <p>In this first investigative study, we examined the factor structure of FFMQ (15 items) using optimal techniques among university students in Pakistan. In addition, we provided a cultural interpretation of the results and conducted a cross‐cultural study while comparing our data with the Spanish sample. It must be noted that FFMQ is an empirically derived scale, and a change in its structure is likely (Baer, [<reflink idref="bib3" id="ref186">3</reflink>]). Our study demonstrated that two factors' higher‐order structure: SR and OE, underpinned by four facets (Observe, Nonreact, Actaware, and Nonjudge) without Describe sub facet, provide a more parsimonious and culturally acceptable account of the FFMQ (15 items) among nonmeditators student sample. The two higher‐order factors were negatively correlated, but each showed strong reliability (>0.8). No difference was found between males and females in mindfulness except for the Nonreact facet. In the first MGCFA, we also achieved metric‐level invariance across gender. In cross‐cultural analysis, our data showed a nonsignificant difference in mindfulness between the Pakistan and Spanish sample barring the Nonjudge facet. However, in the second MGCFA, we only achieved a metric invariance in the correlated model. Our results further suggest that the higher‐order factors operate independently, and the total score cannot be computed. Moreover, in line with Tran et al. ([<reflink idref="bib90" id="ref187">90</reflink>]), we surmise that the Nonreact facet requires revision to include culturally acceptable items. In line with Karl et al. ([<reflink idref="bib56" id="ref188">56</reflink>]), we conjecture that the FFMQ (15 items) is culturally more individualistic, and the inclusion of culturally acceptable items might improve its widespread use. The current study needs to be weighed with certain limitations.</p> <p>In contrast to the original FFMQ framework of a single higher‐order structure, we have presented the theory‐based two‐factor higher‐order structure. However, it requires further evaluation to provide generalizability regarding age, meditation experience, and demographics. Though we have provided theoretical and cultural reasons for removing the Describe facet in the current study, it requires further validation in other samples. The extant literature advocates a few more mindfulness dimensions for collectivist societies, and a newer version of FFMQ is available, such as BIMS (27 items; Padmanabham et al., [<reflink idref="bib75" id="ref189">75</reflink>]). Knowing how newer dimensions and versions might act in Pakistan will be interesting. Another limitation is using a self‐administered questionnaire to a single student sample, which can obscure valid results due to a lack of understanding of items, social desirability, and motivation. Longitudinal studies with different samples might better understand how FFMQ responds. Further cross‐cultural research is required to verify current observations.</p> <hd id="AN0164066401-20">ACKNOWLEDGMENTS</hd> <p>The authors are grateful to Dr. Juan V. Luciano for providing us with the Spanish Pilgrims data. This study was partially funded by the University of Karachi.</p> <hd id="AN0164066401-21">CONFLICT OF INTEREST STATEMENT</hd> <p>The authors declare no conflict of interest.</p> <hd id="AN0164066401-22">DATA AVAILABILITY STATEMENT</hd> <p>The data that support the findings of this study are available from the corresponding author upon reasonable request.</p> <hd id="AN0164066401-23">ETHICS STATEMENT</hd> <p>The Department of Psychology (University of Karachi) approved the study.</p> <ref id="AN0164066401-24"> <title> REFERENCES </title> <blist> <bibl id="bib1" idref="ref56" type="bt">1</bibl> <bibtext> Aguado, J., Luciano, J. V., Cebolla, A., Serrano‐Blanco, A., Soler, J., & García‐Campayo, J. (2015). Bifactor analysis and construct validity of the five facet mindfulness questionnaire (FFMQ) in non‐clinical Spanish samples. 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Header DbId: eric
DbLabel: ERIC
An: EJ1379649
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PubType: Academic Journal
PubTypeId: academicJournal
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Items – Name: Title
  Label: Title
  Group: Ti
  Data: Factor Structure of the Five Facets Mindfulness Questionnaire (FFMQ) (15 Items) in a Collectivist Society--Pakistan
– Name: Language
  Label: Language
  Group: Lang
  Data: English
– Name: Author
  Label: Authors
  Group: Au
  Data: <searchLink fieldCode="AR" term="%22Iqbal%2C+Farhan%22">Iqbal, Farhan</searchLink> (ORCID <externalLink term="http://orcid.org/0000-0002-6227-1979">0000-0002-6227-1979</externalLink>)<br /><searchLink fieldCode="AR" term="%22Iqbal%2C+Farah%22">Iqbal, Farah</searchLink><br /><searchLink fieldCode="AR" term="%22Humayun%2C+Ghazal+Khawaja%22">Humayun, Ghazal Khawaja</searchLink>
– Name: TitleSource
  Label: Source
  Group: Src
  Data: <searchLink fieldCode="SO" term="%22Psychology+in+the+Schools%22"><i>Psychology in the Schools</i></searchLink>. Jul 2023 60(7):2502-2519.
– Name: Avail
  Label: Availability
  Group: Avail
  Data: Wiley. Available from: John Wiley & Sons, Inc. 111 River Street, Hoboken, NJ 07030. Tel: 800-835-6770; e-mail: cs-journals@wiley.com; Web site: https://www.wiley.com/en-us
– Name: PeerReviewed
  Label: Peer Reviewed
  Group: SrcInfo
  Data: Y
– Name: Pages
  Label: Page Count
  Group: Src
  Data: 18
– Name: DatePubCY
  Label: Publication Date
  Group: Date
  Data: 2023
– Name: TypeDocument
  Label: Document Type
  Group: TypDoc
  Data: Journal Articles<br />Reports - Research
– Name: Audience
  Label: Education Level
  Group: Audnce
  Data: <searchLink fieldCode="EL" term="%22Higher+Education%22">Higher Education</searchLink><br /><searchLink fieldCode="EL" term="%22Postsecondary+Education%22">Postsecondary Education</searchLink>
– Name: Subject
  Label: Descriptors
  Group: Su
  Data: <searchLink fieldCode="DE" term="%22Questionnaires%22">Questionnaires</searchLink><br /><searchLink fieldCode="DE" term="%22Metacognition%22">Metacognition</searchLink><br /><searchLink fieldCode="DE" term="%22Personality+Measures%22">Personality Measures</searchLink><br /><searchLink fieldCode="DE" term="%22Foreign+Countries%22">Foreign Countries</searchLink><br /><searchLink fieldCode="DE" term="%22Factor+Structure%22">Factor Structure</searchLink><br /><searchLink fieldCode="DE" term="%22Factor+Analysis%22">Factor Analysis</searchLink><br /><searchLink fieldCode="DE" term="%22College+Students%22">College Students</searchLink><br /><searchLink fieldCode="DE" term="%22Collectivism%22">Collectivism</searchLink><br /><searchLink fieldCode="DE" term="%22Correlation%22">Correlation</searchLink><br /><searchLink fieldCode="DE" term="%22Multivariate+Analysis%22">Multivariate Analysis</searchLink><br /><searchLink fieldCode="DE" term="%22Scores%22">Scores</searchLink><br /><searchLink fieldCode="DE" term="%22Computation%22">Computation</searchLink>
– Name: Subject
  Label: Geographic Terms
  Group: Su
  Data: <searchLink fieldCode="DE" term="%22Pakistan%22">Pakistan</searchLink>
– Name: SubjectThesaurus
  Label: Assessment and Survey Identifiers
  Group: Su
  Data: <searchLink fieldCode="SU" term="%22NEO+Five+Factor+Inventory%22">NEO Five Factor Inventory</searchLink>
– Name: DOI
  Label: DOI
  Group: ID
  Data: 10.1002/pits.22875
– Name: ISSN
  Label: ISSN
  Group: ISSN
  Data: 0033-3085<br />1520-6807
– Name: Abstract
  Label: Abstract
  Group: Ab
  Data: Despite several theoretical, structural, and statistical issues reported against Five Facets Mindfulness Questionnaire (FFMQ), most studies in the educational sector of Pakistan use it without analyzing its factor structure. Since culture might change the structure, this first systematic study filled the gap and explored the factor structure of the least researched FFMQ (15 items) among nonmeditator university students in collectivist society Pakistan while employing optimal statistical techniques. We interpreted our results culturally and also conducted a cross-cultural examination with the Spanish sample. We employed polychoric data to conduct exploratory factor analysis (EFA) and confirmatory factor analysis (CFA). The EFA extracted a four-factor (10 items) solution without the Describe facet and showed good inter-item correlation. The CFA exhibited a theory-based two-factor hierarchical model (self-regulation, orient to experience) with good fit indices, acceptable factor loadings, correct directions, and strong reliabilities (coefficient H > 0.8). The multivariate analysis showed no difference in mindfulness between males and females, except for Nonreact. The first multigroup CFA (MGCFA) showed metric invariance while examining the polarity of items across gender. In cross-cultural multivariate analysis, we found no difference in mindfulness between Pakistan and Spanish samples, except for Nonjudge, but the second MGCFA showed metric invariance in the correlated model only. Our data suggest that the two-factor hierarchical structure, without Describe facet, provides a more parsimonious and culturally accepted structure among nonmeditators in Pakistan. However, due to the independent nature of the two factors, the total score cannot be computed. The FFMQ (15 items) appears individualistic, and including culturally acceptable items will improve its universality.
– Name: AbstractInfo
  Label: Abstractor
  Group: Ab
  Data: As Provided
– Name: DateEntry
  Label: Entry Date
  Group: Date
  Data: 2023
– Name: AN
  Label: Accession Number
  Group: ID
  Data: EJ1379649
PLink https://search.ebscohost.com/login.aspx?direct=true&site=eds-live&db=eric&AN=EJ1379649
RecordInfo BibRecord:
  BibEntity:
    Identifiers:
      – Type: doi
        Value: 10.1002/pits.22875
    Languages:
      – Text: English
    PhysicalDescription:
      Pagination:
        PageCount: 18
        StartPage: 2502
    Subjects:
      – SubjectFull: Questionnaires
        Type: general
      – SubjectFull: Metacognition
        Type: general
      – SubjectFull: Personality Measures
        Type: general
      – SubjectFull: Foreign Countries
        Type: general
      – SubjectFull: Factor Structure
        Type: general
      – SubjectFull: Factor Analysis
        Type: general
      – SubjectFull: College Students
        Type: general
      – SubjectFull: Collectivism
        Type: general
      – SubjectFull: Correlation
        Type: general
      – SubjectFull: Multivariate Analysis
        Type: general
      – SubjectFull: Scores
        Type: general
      – SubjectFull: Computation
        Type: general
      – SubjectFull: Pakistan
        Type: general
      – SubjectFull: NEO Five Factor Inventory
        Type: general
    Titles:
      – TitleFull: Factor Structure of the Five Facets Mindfulness Questionnaire (FFMQ) (15 Items) in a Collectivist Society--Pakistan
        Type: main
  BibRelationships:
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          Name:
            NameFull: Iqbal, Farhan
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            NameFull: Iqbal, Farah
      – PersonEntity:
          Name:
            NameFull: Humayun, Ghazal Khawaja
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          Dates:
            – D: 01
              M: 07
              Type: published
              Y: 2023
          Identifiers:
            – Type: issn-print
              Value: 0033-3085
            – Type: issn-electronic
              Value: 1520-6807
          Numbering:
            – Type: volume
              Value: 60
            – Type: issue
              Value: 7
          Titles:
            – TitleFull: Psychology in the Schools
              Type: main
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