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Psychometric Evaluation of the Identification with the Country Scale in a Chilean Sample

A peer-reviewed version of this preprint was published in:
Behavioral Sciences 2026, 16(7), 1106. https://doi.org/10.3390/bs16071106

Submitted:

09 June 2026

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11 June 2026

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Abstract
Group identification refers to a psychological connection with a group whereby individuals incorporate group-defining characteristics into their self-concept. The scale developed by Mael & Ashforth is one of the most used, but for knowledge, it has not been studied in Chile. This study analyzed the psychometric properties of this scale in a Chilean sample using the country as the reference group. We expected a one-factor structure and convergent validity. In a non-probabilistic sample (N = 511), we evaluated the one-factor fit of the scale throug. h confirmatory factor analysis. Results indicated an acceptable fit (χ2/df = 34.232/9 = 3.80, p < .001, CFI = .974, TLI = .956, SRMR = .031, and RMSEA = .089). Likewise, factor loadings were adequate (> .592), and the model presented a high reliability (? > .870). Also, convergent validity was found, showing that identification with Chile and identity fusion with Chile measurements are related (r = .50, p < .001), although they represent different constructs. Finally, the model showed invariance for gender. The relevance of the scale was discussed according to the possible positive and negative effects of identification with the country, in particular in contexts with an important migration process.
Keywords: 
group identification
;  
social identity theory
;  
factor structure
;  
identity fusion
Subject: 
Social Sciences  -   Psychology

1. Introduction

One of the foundational frameworks in social psychology is social identity theory, which seeks to explain how people behave in terms of their social, rather than personal, identities (Ellemers & Haslam, 2012). This theory posits that, in many scenarios, individuals categorize themselves and others into social groups in order to enhance self-esteem (Tajfel, 1974; Tajfel & Turner, 1979; Turner & Reynolds, 2012). Subsequent work has suggested that self-esteem is not the sole motive explaining that categorization (Ellemers & Haslam, 2012). Social identity theory further outlines the psychological dynamics involved in the distinction between personal and social identity, the strategies individuals use to achieve a positive social identity, and the structural characteristics of the social context that influence which strategies are more likely to be adopted in a given situation (Ellemers & Haslam, 2012).
Personal identity refers to a part of self-concept composed of idiosyncratic characteristics (e.g., individual abilities), whereas social identity denotes the component of the self-concept derived from group membership and the emotional significance attached to that membership (Tajfel, 1974; Turner & Reynolds, 2012). A central concept in social identity theory that links both personal and social identity is group identification. This refers to a psychological connection with a group whereby individuals incorporate group-defining characteristics into their self-concept, such that social identity becomes salient and predominates over personal identity (Mael & Ashforth, 1992; Riketta, 2005; Tajfel & Turner, 1979). Within this framework, personal and social identities are often conceptualized as operating in a hydraulic manner; that is, when one identity is salient, the other recedes (Turner & Reynolds, 2012). Likewise, subjects tend to perceive as more relevant those group members who embody prototypical characteristics, whereas other members are viewed as relatively interchangeable (Swann et al., 2012; Yzerbyt & Demoulin, 2010). Moreover, the strength of identification with a given group is dynamic and may vary over time and across social contexts (Doosje et al., 2002).
Group identification is influenced by the status of the group. When people define themselves as members of a high-status group, they are more likely to achieve a positive social identity compared to members of low-status groups (Ellemers, 1993). When such distinctiveness is an integral part of the individual's identity (Spiteri et al., 2026), high group status is associated with feelings of admiration among its members, which in turn enhances self-esteem (Mael & Ashforth, 2001).
Group identification is grounded in individuals’ need to align their sense of self with a collective (Ashforth & Mael, 2024). For example, in organizational contexts, group identification involves feeling pride in group membership and internalizing the group’s values (Riketta, 2005). Consequently, belonging to a group shapes individuals’ attitudes and behavior, particularly when they define themselves in terms of their group membership (Van Knippenberg & Van Schie, 2000). Accordingly, group identification promotes a range of intragroup behaviors that strengthen the group (Tajfel & Turner, 1979; Varmann et al., 2023; Zhang et al., 2022), such as increased work engagement (Rongzhi & Qing, 2026), organizational commitment (Van Knippenberg & Sleebos, 2006), reduced turnover intentions (Basar & Filizoz, 2015), and greater group cohesion, as observed among active runners (Stevens et al., 2019) and among police officers and firefighters (García-Guiu et al., 2015). In addition, group identification is associated with positive outcomes beyond the group context, including work satisfaction across occupations (Lee et al., 2015; Ng, 2015; van Knippenberg & van Schie, 2000), lower levels of teachers’ burnout (Uzun, 2018), and increased engagement in collective action to support individuals with disabilities (Pérez-Garín et al., 2021). Moreover, identification with the country, in particular, can be linked to both positive and negative outcomes. On one hand, it has been associated with increased well-being over time (Khan et al., 2020) and adaptive responses in stressful contexts such as the COVID-19 pandemic (Bonetto et al., 2022). On the other hand, it may also be associated with negative consequences, including prejudice among locals toward immigrants (Author et al., 2024) and reciprocal prejudice among immigrant populations (González et al., 2010).

1.1. Group Identification Measurements

A variety of instruments have been used to assess group identification. Most of these instruments consist of Likert-type scales in which individuals report their level of agreement with items referring to their relationship with a specific group or social entity. In the present study, we focused on the scale developed by Mael and Ashforth, as it is one of the most widely used measures of identification (Riketta, 2005), and has demonstrated a stronger prediction power for extreme pro-group behaviors compared to other instruments, such as those proposed by Ellemers et al. (1999), Leach et al. (2008), Postmes et al. (2013), and Steffens et al. (2015; for a review, see Varmann et al., 2023). Originally developed in organizational settings, Mael and Ashforth’s scale has been successfully extended to a variety of contexts, including identification with different types of groups, such as national identity (Gómez et al., 2011), religious groups (Besta et al., 2014), and even non-social entities such as store brands (Rubio et al., 2015).
The Mael and Ashforth´s scale is a self-report Likert-type of instrument that has consistently shown a unidimensional (one-factor) structure across diverse cultural contexts including samples from United States (Mael & Ashforth, 1992), Turkish (Δχ2/sd =1.05, GFI = .98, NFI = .99, RMSA = .02, Basar & Filizoz, 2015; χ2/sd = 1.70, CFI = .99, NFI = .99, GFI = .99, AGFI = .98, TLI = .99, RMSEA= .04, SRMR = .02, Öz & Meriç, 2026), Chinese (χ2/df = 2.696, RMSEA = .073, CFI = .993, TLI = .987, and SRMR = .012, Yang et al., 2022; Rongzhi & Qing, 2026), and Romanian (Boroş, 2008) samples. Reported reliability has been acceptable α > .70 across studies conducted in different countries (Contreras-Pacheco & Lesmez-Peralta, 2018; García-Guiu et al., 2015; Irshad & Bashir, 2020; Pérez-Garín et al., 2021; Rodríguez-Carvajal et al., 2014; van Knippenberg & van Schie, 2000). Additionally, factor loadings have typically been adequate > .50 in samples from the United States (Anaza, 2015), Spain (García-Guiu et al., 2015; Gómez et al., 2011), Turkey (Basar & Filizoz, 2015; Öz & Meriç, 2026), Colombia (Contreras-Pacheco & Lesmez-Peralta, 2018), and Romania (Boroş, 2008).

1.2. The present Study

As noted above, the Mael and Ashforth scale has been applied to a wide range of referent targets, including organizations, social groups, and abstract entities. In the present study, we focus on identification with the country as the target. Although this construct has been extensively examined in developed societies, it has received comparatively less attention in Latin American contexts. To our knowledge, the psychometric properties of Mael and Ashforth's scale have not yet been examined in a Chilean sample. Accordingly, this study aims to evaluate the psychometric properties of the Mael and Ashforth scale in a Chilean sample.
We proposed the following hypothesis:
H1: The scale will show structural validity consisted in a one-factor model, with acceptable factor loadings (> .500) and fit indexes (χ2/df < 5; CFI > .95; TLI > .95; RMSEA < .08, SRMR < .08).
H2: Identification with Chile will show adequate reliability (α >.70).
Convergent validity was assessed using identity fusion, a construct that captures a strong relational bond with a group. Identity fusion refers to a visceral sense or feeling of oneness with a group that motivates pro-group behaviors, including a willingness to engage in extreme actions or personal sacrifice for the benefit of other group members (Gómez et al., 2020; Swann et al., 2015). A substantial body of research has shown that identity fusion robustly explains pro-group behaviors (Buhrmester et al., 2014; Gómez, Brooks et al., 2011; Swann, Buhrmester et al., 2014; Swann et al., 2010; Swann et al., 2012; Whitehouse et al., 2014) and differs from traditional group identification in several key respects. First, identity fusion is characterized by identity synergy, whereby personal and social identities are experienced as functionally equivalent and mutually reinforcing. In contrast, models of group identification conceptualize these identities as operating in a compensatory or a “hydraulic” manner (Turner & Reynolds, 2012). Second, identity fusion involves strong relational ties, fostering deep, familial-like bonds among group members, in which members are valued for their unique personal qualities. By contrast, group identification emphasizes prototypicality, whereby individuals who best represent the group's defining characteristics are more highly valued (Swann et al., 2012; Yzerbyt & Demoulin, 2010). Third, identity fusion tends to be more stable and enduring over time, whereas group identification is more context-dependent and variable (Gómez et al., 2020).
Although identification with the country and identity fusion with the country are positively associated (.29 < r < .88; Besta et al., 2014; Gómez et al., 2011), factor analytic evidence has shown that they are empirically different constructs (Gómez et al., 2011). According to this literature, we hypothesized:
H3: Scores on the identification scale will be positively associated with identity fusion.
This research provides evidence for the psychometric properties of the scale, a necessary condition for its rigorous application in future research, in which group identification may play a central explanatory role.

2. Materials and Methods

The present study employed a cross-sectional instrumental design (Ato et al., 2013). We examined the psychometric properties of Mael and Ashforth's group identification scale in a Chilean sample. This study was conducted as part of a broader research project aimed at evaluating the effect of identity fusion with the country in predicting prejudice toward immigrants.

2.1. Participants

A non-probabilistic sample of 511 Chilean participants completed this study via online survey (age range between 18 and 59 years old, M = 22.8, SD = 7.00; 60.86% female). Participants reported diverse occupational positions, including students (61.25%), students and workers (26.42%), and workers (12.33%). Monte Carlo simulations showed that this sample size with error type 1 (.05), lets to detect factor loadings > .500 with a power < .99, which is higher than the recommended value of .80.

2.2. Instruments

2.2.1. Identification with the Country

We adapted the group identification scale developed by Mael and Ashforth (1992) using a back-translation procedure (Berry, 1980), replacing references to “school” (or the specific institution) with “my country”. Participants indicated their agreement with six statements assessing their relationship with their country on a Likert-type scale ranging from 0 (Totally disagree) to 4 (Totally agree). Example items include “When someone criticizes my country, it feels like a personal insult” and “The successes of my country are my successes”. Previous research has reported good reliability for this scale (α between.74 and .90; García-Guiu et al., 2015; Gómez et al., 2011; Jones & Volpe, 2011; Rongzhi & Qing, 2026; Ozturk & Soyturk, 2021).

2.2.2. Identity Fusion with the Country

Identity fusion was assessed using the Spanish version of the verbal fusion scale (Gómez et al., 2011). Participants rated their agreement with seven items describing their relationship with their country on a scale ranging from 0 (strongly disagree) to 6 (strongly agree). Example items include “I am one with my country,” and “I have a deep emotional bond with my country” (αs ranging from .70 to .95; Besta et al., 2014, 2015; Buhrmester et al., 2014; Gómez et al., 2011; Author et al., 2022; Swann, Buhrmester et al., 2014; Swann, Gómez et al., 2014). In the present study, the reliability was very high, α = .92.

2.3. Procedure

Participants were recruited through an online survey distributed via social media (e.g., Facebook and Instagram) and professional networking (e.g., LinkedIn) platforms, institutional emailing lists from Chilean universities, and personal networks. Recipients were also encouraged to share the survey with their personal contacts, thereby increasing sample reach.
The invitation included a link to the study, which participants accessed voluntarily. Upon accessing the survey, participants provided informed consent electronically before proceeding. Then they completed the identity fusion and group identification scales and provided their demographic information. The survey required approximately four minutes to complete. Responses with missing data were excluded from the analysis.

2.4. Data Analysis

We conducted a confirmatory factor analysis to evaluate the factor structure of the scale and assessed reliability using Cronbach's alpha. Factor loading > .500 were considered acceptable (Hair et al., 1999). Given the sensitivity of the χ2 statistic to sample size, particularly when N > 200, where well-fitting models are often rejected (Hair et al., 1999; Ruiz et al., 2010). Model fit was evaluated using multiple indices. Specifically, acceptable model fit was defined as χ2/df < 5 (Kline, 2016), TLI > .95, CFI > .95, RMSEA < .08, and SRMR < .08 (Mîndrilă, 2010; Schreiber, 2017). Reliability was interpreted according to the following criteria: very low (α < .30), low (.30 < α ≤ .60), moderate (.60 < α ≤ .75), high (.75 < α ≤ .90), and very high (α > .90; Alexandre et al., 2013)
Convergent validity was assessed by examining the correlation between the identification scale and the identity fusion measure. To further evaluate the distinction between both constructs, alternative measurement models were compared, considering differences to be meaningful when |ΔCFI| > .010 (Cheung & Rensvold, 2002). Measurement invariance across gender was also examined to determine whether the identification scale operates equivalently for women and men (coded as 0 and 1, respectively). Configural, metric, and scalar invariance models were tested sequentially. Following Chen (2007), invariance was supported when changes in fit indices met the following criteria to support invariance: for metric invariance (|ΔCFI| < .010, |ΔRMSEA| < .015, and |ΔSRMR| < .030; for scalar and strict invariance (|ΔCFI| < .010 and |ΔRMSEA| < .015, and |ΔSRMR| < .010). All analyses were conducted using the software RStudio and the Lavaan Package (Rosseel, 2012).

2.5. Ethical Aspects

This study was conducted as part of a broader project examining predictors of prejudice towards outgroups. The project, which included the data used in this study, was approved by the Ethical Committee of X University (anonymity) and carried out in accordance with institutional guidelines and the principles of the Declaration of Helsinki. Participation was voluntary. Participants provided their informed consent electronically before completing the survey and then responded to the scales detailed above.

3. Results

Descriptive statistics and inter-item correlations are presented in Table 1. Multivariate normality was assessed using Mardia’s test (Mardia, 1974), which indicated significant departures from normality (skewness: b = 2.44, p < .001; kurtosis: b = 57.04, p < .001). Accordingly, Spearman’s rank-order correlations were computed, and confirmatory factor analysis was conducted using a robust maximum likelihood estimation (RMLE), which is appropriate for non-normal data (Muthén & Muthén, 1998-2017). All items were positively and significantly correlated (p < .001).
The unidimensional model demonstrated an overall acceptable fit to the data (χ2/df = 34.232/ 9 = 3.80, χ2 was significant at p < .001, CFI = .974, TLI = .956, RMSEA = .089, and SRMR = .031). Although the RMSEA value slightly exceeded the conventional cut-off (RMSEA < .080), this index is known to be biased upward in models with low degrees of freedom and may overestimate model misfit in such cases (Kenny et al., 2015). Importantly, the remaining fit indices indicated good model fit. All factor loadings were adequate (> .500, see Table 2), supporting the unidimensional structure of the scale. In addition, the scale demonstrated high reliability, α = .87.
Convergent validity was examined by assessing the relationship between identification with Chile and identity fusion with Chile. As expected, both constructs were positively related, r = .50, p < .001. To further evaluate their distinctiveness, we compared a one-factor model with a two-factor model. The two-factor model showed a better fit to the data (χ2/df = 5.33, CFI = .909, TLI = .889, RMSEA= .102, and SRMR = .052) than the one-factor model (χ2/sd = 15.11, CFI = .717, TLI = .660, RMSEA= .166, and SRMR = .123), with the difference in fit indices exceeding the recommended threshold |ΔCFI| > .010.
Finally, measurement invariance across gender was evaluated (see Table 3). The configural model demonstrated acceptable fit (CFI > .95, TLI > .95, RMSEA < .08, and SRMR < .08), indicating a similar factor structure across groups. Subsequent comparisons showed that the differences between configural and metric models, metric and scalar models, and scalar and strict models met the recommended criteria |ΔCFI| < .010 and |ΔRMSEA| < .015, and particularly, |ΔSRMR| < .030 for metric, and |ΔSRMR| < .010 for scalar and strict models.

4. Discussion

This study examined the psychometric properties of the identification with the country scale developed by Mael and Ashforth (1992) in a Chilean sample. The results indicated a good fit for a one-factor structure, with adequate factor loadings and reliability. As expected, identification with the country was positively associated with identity fusion with the country. These findings add to the extensive body of literature supporting the validity of this group identification scale across samples of different countries (e.g., Basar & Filizoz, 2015; Boroş, 2008; Rongzhi & Qing, 2026; Mael & Ashforth, 1992; Yang et al., 2022).
Assessing identification with the country may be particularly relevant in contexts such as Chile, as it can help to better predict a range of social and psychosocial outcomes. On the one hand, in developing countries, challenges related to quality of life may be mitigated with the increased well-being associated with national identification (Khan et al., 2020), especially in adverse scenarios (Bonetto et al., 2022). On the other hand, group identification may also be associated with negative outcomes, for example, studies in Chile have reported prejudice among locals toward immigrants (Author et al., 2024) as well as prejudice among immigrants toward natives (González et al., 2010). This issue is particularly important in countries such as Chile, which have experienced substantial increases in immigration in recent years (Instituto Nacional de Estadísticas & Servicio Nacional de Migraciones, 2024).
The unidimensional structure, adequate reliability (> .70), and factor loadings (> .500) observed in this study are consistent with those reported in previous research (e.g., Anaza, 2015; Basar & Filizoz, 2015; Contreras-Pacheco & Lesmez-Peralta, 2018; García-Guiu et al., 2015; Gómez et al., 2011; Jones & Volpe, 2011; Mael & Ashforth, 1992; Rongzhi & Qing, 2026). The consistency of these findings across studies suggests that the Mael and Ashforth scale is a robust instrument that remains stable across different cultural contexts. Future research could extend these findings by examining predictive and longitudinal validity, which, to our knowledge, has not yet been explored in the Chilean context.
Identification with the country was positively associated with identity fusion with the country, supporting the convergent validity of the scale and aligning with previous research (Besta et al., 2014; Gómez et al., 2011; Swann et al., 2010). Likewise, factor analysis indicated that these constructs are different, as reported in prior studies (Gómez et al., 2011). This pattern may be explained by the fact that both constructs reflect forms of group bonding, yet differ in their underlying properties, as it was explained before (for a review, see Gómez et al., 2020).
Measurement invariance analysis showed that the scale functions equivalently across men and women. Thus, the instrument appears to assess the construct in the same way across genders. Hence, future studies can compare both genders, and any difference found would not be attributable to measurement bias, but rather to actual differences in group identification.
The limitations of this study relate to sampling and the choice of reference group. First, the use of a non-probabilistic sample limits the generalizability of the results to the broader Chilean population. Second, the psychometric properties of the scale were assessed using the country as the reference group, and thus, the results may not generalize to other types of groups. This is consistent with the notion that social identification is not a stable dispositional trait (Weber et al., 2011) but rather depends on contextual factors such as group distinctiveness (Grant & Hogg, 2012) and entitativity (Jans et al., 2011). Despite these limitations, the present findings provide initial evidence supporting the reliability and validity of the identification with the country scale in a Chilean context. These results contribute to the cross-cultural literature on group identification and highlight the importance of considering both its positive and negative implications. Future research should continue to examine this construct across diverse populations and contexts to further refine its theoretical and empirical understanding.

Institutional Review Board Statement

The study was conducted in accordance with the Declaration of Helsinki, and approved by the Comité Ético Científico de la Universidad de Atacama (Approval Code: ORD. N°17/2024 (Original) / ORD. N°81/2024 (Amendment), Approval Date: 15 April 2024 (Original) / 18 November 2024 (Amendment)).

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Table 1. Descriptive statistics and correlations among items.
Table 1. Descriptive statistics and correlations among items.
Mean (SD)
Correlations
1 2 3 4 5
Item 1. When someone criticizes my country, it feels like a personal insult. 2.19 (1.39)
Item 2. I am very interested in what citizens of other countries think about my country. 1.80 (1.35) .50
Item 3. When I talk about my country, I usually say “we” rather than “they.” 2.68 (1.35) .45 .40
Item 4. The successes of my country are my successes. 1.95 (1.37) .60 .45 .57
Item 5. When someone praises my country, it feels like a personal compliment. 2.11 (1.40) .68 .49 .52 .74
Item 6. If a story in the media criticized my country, I would feel embarrassed. 1.96 (.137) .49 .44 .45 .48 .56
Table 2. Factor loadings of the identification with Chile scale items.
Table 2. Factor loadings of the identification with Chile scale items.
Item Factor loading
1 .764***
2 .592***
3 .648***
4 .815***
5 .875***
6 .640***
Note. ***p < .001.
Table 3. Configurational, metric, scalar, and strict models for gender invariance.
Table 3. Configurational, metric, scalar, and strict models for gender invariance.
Model CFI ΔCFI TLI RMSEA ΔRMSEA SRMR |ΔSRMR|
Configural .977 .961 .068 .029
Metric .968 .009 .958 .071 .003 .055 .026
Scalar .964 .004 .961 .068 .003 .058 .003
Strict .961 .003 .965 .065 .003 .054 .004
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