Assessing Individual Variation in Action Identification: Translation,Validation,and Shortening of the Polish Behavior Identification Form (BIF)

Abstract

Action identification theory asserts that level of action identification holds implications for important aspects of personality, including self-regulation, self-concept, and vulnerability to social influence. Vallacher and Wegner’s Behavior Identification Form is a 25-item forced-choice instrument that assesses individual differences in characteristic level of action identification and is widely used in social and psychological research. This article presents studies that evaluated the psychometric properties of the Polish version of the Behavior Identification Form, further modified it into a shortened version, and investigated several predicted conceptual relationships between the Behavior Identification Form and other personality measures. A total of 1153 Polish university students in three samples took part in the research. Psychometrically, the Polish Behavior Identification Form proved to be reasonably sound, with good internal consistency (avg. α = .83 and α = .72 for the full and short form, respectively). The predictive abilities of the full and short Polish Behavior Identification Form were consistently comparable, although not all of the expected associations were found.

Keywords

Action identification personal agency Behavior Identification Form translated questionnaires psychometrics self-regulation self-understanding

Introduction

The aim of the present investigation was to develop a Polish language version of an individual difference measure of action identification, based on translating and refining the Behavior Identification Form (BIF), developed originally by Vallacher and Wegner (1989). The concept of action identification links with Pennebaker’s (1989) levels of thinking construct, Carver and Scheier’s (1982) notion of hierarchical levels of control, and Trope and Liberman’s (2010) concept of level of construal and is essentially about providing a framework for understanding an array of issues that center on mental control of action. It has also proven to be an important dimension along which individuals differ.

Definitions

Action identification theory (Vallacher & Wegner, 1987, 1989) holds that people know what they are doing and that they identify their actions on multiple hierarchical levels, ranging from low-level identities that specify how the action is performed (e.g., mopping the floor) to high-level identities that signify why or with what effect the action is performed (e.g., getting ready for company). At a particular point in time, only one identity from the individual’s act identity structure tends to be prepotent, reflecting a trade-off between comprehensive understanding and effective performance of the action. The prepotent identity serves as a cognitive referent for initiating, carrying out, and evaluating the action’s performance. People tend to appreciate high-level identities, because of comprehensive meaning of an action that is provided by such identities. However, when an action cannot be performed effectively under a given identity, people are likely to embrace a lower level identity in order to figure out how to continue. Hence, the optimal level of action identification matches the individual’s capacity to perform the act. Both the identification of a personally easy action in relatively low-level terms and the identification of a difficult or unfamiliar action in high-level terms illustrate a nonoptimal action identification, which can lead to performance impairment and heightened anxiety and self-focused attention (Vallacher, 2007).

Level of action identification tends to vary across situations (action-specific variation). Nevertheless, people differ in their tendencies to understand their actions in terms of their consequences and implications versus in terms of their details or mechanics, and these differences are relatively stable, representing people’s standing on a broad personality dimension—level of personal agency. Part of this individual variation in identification level across action domains may stem from differences in competence as well as in the frequency with which individuals encounter life situations that remind them of the higher level meanings of their actions (Vallacher & Wegner, 1987, 1989).

Implications of level of personal agency

There is evidence suggesting that individual differences in level of action identification have implications for the individual’s overall competence in action and inclination toward planful versus impulsive action as well as for the degree to which the individual’s actions are organized by and reflected in the self-concept. More specifically, compared with low-level agents, high-level agents judge various everyday actions as less difficult, claim greater proficiency at wide array of activities, are less prone to chronic disruption in their actions, report lesser impulsiveness, score more internal in locus of control, and show greater persistency and stability in their actions across contextual variations (e.g., Fujita, Trope, Liberman, & Levin-Sagi, 2006; Vallacher & Wegner, 1989). Moreover, those controlling their actions at a relatively high-level display more attitude–intention congruence and have a more clear and stable sense of what they are like than individuals who identify their actions at lower levels (e.g., McIntyre, Paulson, Lord, & Lepper, 2004; Vallacher & Wegner, 1989). High-level agents also tend to be more open to experience and less neurotic (Levesque, 2012).

Development of original (English) scale

The original BIF was developed by Vallacher and Wegner (1989). It is a 25-item inventory that lists midlevel neutral actions (e.g., “making a list”) along with alternative identifications: one lower in level, describing the behavior in terms of its means (e.g., “writing things down”) and one higher in level, describing it in terms of its reasons, effects, or implications (e.g., “getting organized”), and asks participants to choose which one of two descriptions they “personally believe is more appropriate in each pair.” Total scores are based on the number of high-level alternatives chosen, with higher scores indicating stronger preferences for high-level action identification.

The alternate identifications were derived from the redescriptions of various specific behaviors most frequently mentioned by pilot participants in an open-ended pretesting. An initial sample of 60 items was subjected to mass testing and statistical analysis, eliminating items that had low item–total correlations. The long list was reduced to 25 items, which showed good internal consistency (Cronbach’s α of .85) and test–retest reliability over a two-week period (r_tt = .91). An exploratory factor analysis (EFA), using a principal axis solution, indicated one primary factor (eigenvalue = 5.65) and four minor factors. However, the varimax rotation failed to converge after 24 iterations, suggesting that the BIF was a one-dimensional scale (Vallacher & Wegner, 1989).

Overview

The studies presented here were conducted with the aim of developing and validating a Polish version of the BIF. The first study aimed to translate the BIF into Polish, determine the equivalence of the Polish version with the English BIF, and establish the preliminary psychometrics of the translated instrument with a bilingual sample. The second and third studies were conducted to continue the psychometric evaluation of the Polish version of the BIF. They focused on confirming its reliability and validity. In order to examine the structural validity of the BIF, EFA and confirmatory factor analysis (CFA) were used.

Criterion validity was examined by estimating correlations with measures of self-regulation (i.e., self-control, cognitive failures, and locus of control), goal orientation (i.e., goal concordance and goal coherence), and self-understanding (i.e., self-concept clarity, sense of identity, and self-esteem). These correlates were chosen to reflect various features of personality functioning and allow for comparison with previous studies, including the original study of Vallacher and Wegner (1989). According to action identification theory, familiar and routine tasks are best maintained with respect to relatively high-level identities (Vallacher & Wegner, 1989). It was thus expected that low-level agents would be more prone than high-level agents to disruption in their everyday actions. Theoretical evidence also suggests that individual variation in action identification implies individual variation in premeditation of behavior, with high-level identification being linked to intrinsic motivation and high effortful self-control (Vallacher & Wegner, 1989). Therefore, high-level agents were expected to be less inclined toward impulsive responding, hold internal control beliefs, and pursue self-concordant and coherent goals to a greater extent than low-level agents. Finally, since high-level identities carry more meaning and importance than low-level identities, high-level agents are likely to have a well-defined self-conception (Vallacher & Wegner, 1989). Consistent with that, it was predicted that individuals with high levels of personal agency would be characterized by a clear and stable self-understanding. Levels of personal agency, however, should be unrelated to self-evaluation (Vallacher & Wegner, 1989).

Furthermore, given that various versions of the BIF have been used in previous studies, with a number of items ranging from 13 (e.g., Fujita, Henderson, Eng, Trope, & Liberman, 2006) to 25 (e.g., Levy, Freitas, & Salovey, 2002), the possibility of shortening the Polish BIF has also been tested. Item selection for the shortened measure was based on a combination of internal, external, and judgmental criteria, as recommended by Stanton, Sinar, Balzer, and Smith (2002).

Method

Participants

A total of 1153 participants in three samples participated in this research. All samples consisted of undergraduate students majoring in different academic disciplines at Poznań (Poland) universities. Table 1 shows each sample’s size, age, and gender distribution.

Table 1.

Samples characteristics.

		Age		Gender
Sample	n	Range	M (SD)	Women n (%)	Men n (%)
Sample 1	84	21–34	23.01 (1.79)	62 (73.8)	22 (26.2)
Sample 2	349	18–29	20.92 (1.53)	215 (61.6)	134 (38.4)
Sample 3^a	720	18–31	20.88 (1.48)	396 (55.1)	323 (44.9)
Total	1153	18–34	21.04 (1.61)	673 (58.4)	479 (41.6)

^aGender information was unavailable for one participant.

Measures

Ten measures were obtained for the studied samples. Each sample completed the Polish translation of the BIF. In addition, Sample 1 completed the English version of the BIF, and the other two samples filled in a set of other measures that were relevant to the personal agency construct. The Self-Control Scale (SCS), the Events Related to Innatention Scale (ERIS), Rotter’s Internal–External (IE) Control Scale, the Personal Striving Questionnaire (PSQ), the Striving Instrumentality Matrix (SIM), the Self-Concept Clarity (SCC) Scale, the Multidimensional Questionnaire of Identity–Extended (MQI-E), and Rosenberg’s Self-Esteem Scale (SES) were used to measure the variables of interest.

Equivalence of the translation

Participants and procedure

Participants in Sample 1 were 84 MA English students of a public university in Poznań (Poland), fluent and literate in English and Polish. They were administered the English and the translated Polish version of the BIF, approximately two weeks apart. The participation was voluntary and based on informed consent. No participation payment was made.

Measures

As described earlier, the original version of the 25-item BIF was prepared in English. The translation process of the BIF followed a rigorous methodology that consisted of (1) three independent forward translations produced by three professional translators; (2) forward translations synthesized into one by three experts (two with PhDs in psychology and one with a PhD in linguistics); (3) three independent blind back-translations produced by three professional translators; (4) back-translations synthesized into one by two experts (with PhDs in psychology); and (5) the consensus back-translated English version reviewed and accepted by one of the authors of the BIF, i.e., R. R. Vallacher. This process resulted in a prefinal Polish version of the BIF for field testing (see Appendix 1 for the final translations of each item; the corresponding original items can be found in Vallacher & Wegner, 1989).

Results

Responses on both versions of the BIF were compared using descriptive statistics, McNemar or Wilcoxon signed-rank test, and correlation coefficients (Table 2). Relatively small differences were found between English items and their Polish counterparts (Cohen’s g effect size ranged from .00 to .30; average g = .11). These differences were significant only for Items 6, 20, and 24. The correlation between total scores on the two language versions was high (r =.80, p < .001), especially if considering that the association took into account two sources of variability: language and time. The difference between total scores on the two versions was not significant (r = .14).

Table 2.

Descriptive statistics and cross-language comparison of item and total scores.

	English version			Polish version			McNemar/Wilcoxon test
Item	M (SD)	r _it	α if item deleted	M (SD)	r _it	α if item deleted	Statistic	Effect size
BIF1	0.49 (0.50)	.25	.83	0.46 (0.50)	.29	.86	0.03	.03
BIF2	0.74 (0.44)	.26	.83	0.71 (0.45)	.23	.86	0.05	.05
BIF3	0.52 (0.50)	.12	.83	0.63 (0.49)	.41	.86	2.23	.18
BIF4	0.44 (0.50)	.32	.83	0.38 (0.49)	.44	.86	0.76	.12
BIF5	0.68 (0.47)	.37	.82	0.75 (0.44)	.55	.85	0.89	.11
BIF6	0.20 (0.40)	.47	.82	0.36 (0.48)	.41	.86	6.50**	.27
BIF7	0.37 (0.49)	.31	.83	0.49 (0.50)	.59	.85	2.70	.17
BIF8	0.75 (0.44)	.23	.83	0.68 (0.47)	.55	.85	1.14	.14
BIF9	0.43 (0.50)	.50	.82	0.54 (0.50)	.42	.86	2.56	.18
BIF10	0.35 (0.48)	.28	.83	0.37 (0.49)	.44	.86	0.04	.04
BIF11	0.08 (0.28)	.47	.82	0.12 (0.33)	.41	.86	0.57	.21
BIF12	0.50 (0.50)	.41	.82	0.55 (0.50)	.45	.86	0.35	.08
BIF13	0.44 (0.50)	.30	.83	0.43 (0.50)	.46	.85	0.00	.02
BIF14	0.76 (0.43)	.54	.82	0.70 (0.46)	.49	.85	0.84	.13
BIF15	0.33 (0.47)	.52	.82	0.40 (0.49)	.44	.86	1.14	.14
BIF16	0.49 (0.50)	.59	.81	0.49 (0.50)	.62	.85	0.00	.00
BIF17	0.27 (0.45)	.42	.82	0.26 (0.44)	.47	.85	0.00	.02
BIF18	0.32 (0.47)	.51	.82	0.39 (0.49)	.50	.85	1.04	.13
BIF19	0.33 (0.47)	.39	.82	0.33 (0.47)	.34	.86	0.00	.02
BIF20	0.35 (0.48)	.49	.82	0.18 (0.39)	.25	.86	7.68**	.32
BIF21	0.31 (0.47)	.32	.83	0.32 (0.47)	.41	.86	0.00	.02
BIF22	0.86 (0.35)	.07	.83	0.80 (0.40)	.05	.87	0.84	.13
BIF23	0.63 (0.49)	.45	.82	0.62 (0.49)	.53	.85	0.00	.02
BIF24	0.58 (0.50)	.44	.82	0.45 (0.50)	.32	.86	4.00*	.22
BIF25	0.68 (0.47)	.26	.83	0.71 (0.45)	.23	.86	0.17	.07
Total	11.21 (5.13)	.37^a	.83^b	11.87 (5.64)	.41^a	.86^b	1.81	.14

BIF: Behavior Identification Form.

^aAverage corrected item–total correlation.

^bCronbach’s alpha coefficient.

*p < .05, **p < .01.

Estimates of internal consistency reliability and homogeneity were similar in both the English and Polish versions of the scale (Table 2). Cronbach’s alphas were .83 and .86 for the English and Polish versions of the BIF, respectively. The majority of item–total correlations were above .30 (r_it ranged from .07 for Item 22 to .59 for Item 16, and from .05 for Item 22 to .62 for Item 16 for the English and Polish BIF, respectively). Item-deleted coefficient alpha statistics indicated that only Item 22 in the Polish BIF would result in an increase in reliability (from .86 to .87) if it was eliminated. EFA was not conducted at this time due to the small sample size of the bilingual sample.

Discussion

Overall, the Polish version of the BIF appeared to be compatible with the original English scale and similarly reliable. Moreover, coefficient alpha and average item–total correlation for the Polish BIF were slightly higher than that of Vallacher and Wegner’s (1989) initial analysis (α = .85, avg. r_it = .35, respectively). The few problematic items generated further discussion. Since the scale already demonstrated good reliability, and considering that some items may have been underrepresented in the study sample, the research team decided to keep all items in the pool.

Factor analysis and reduction of the Polish BIF

Participants and procedure

The factor analyses were conducted on Sample 2, which comprised 349 participants, and Sample 3, which comprised 720 participants. Participants in both samples were students in different faculties in universities in Poznań. Many of them came from the same university, but different academic units than Sample 1. The assessments were conducted during academic class hours by trained researchers. The order in which participants completed the questionnaires was counterbalanced. They were informed that the participation was voluntary and that confidentiality was guaranteed. No financial incentives for completing the questionnaires were offered, but participants from Sample 3 were informed of an opportunity to participate in a follow-up study, for which they would receive a cinema voucher.

Measures

All participants completed the Polish version of the BIF, described above, along with other measures.

Results

First, the internal consistency of the Polish BIF in each sample was assessed. The measure was found to have satisfactory reliability with Cronbach’s α of .84 and .80 for Sample 2 and Sample 3, respectively. As Table 3 indicates, for both samples, the corrected item–total correlations were all positive and not less than the rule of thumb minimum value of .20 (Kline, 1986), except for Items 20 and 22. The item–total correlations for Item 20 were .19 and .13 for Sample 2 and Sample 3, respectively, and the item–total correlations for Item 22 were .18 and .02 for Sample 2 and Sample 3, respectively. However, in both cases, deletion of the items with the lowest item–whole correlation did not result in an increase in the α coefficient. In fact, deletion of none of the items resulted in an increase in α.

Table 3.

Descriptive statistics, reliability information, and EFA and CFA factor loadings.

	Sample 2				Sample 3
Item	M (SD)	r _it	α if item deleted	EFA	M (SD)	r _it	α if item deleted	CFA
BIF1	0.56 (0.50)	.29	.83	.32	0.57 (0.50)	.29	.79	.42
BIF2	0.84 (0.37)	.34	.83	.37	0.88 (0.32)	.28	.79	.50
BIF3	0.60 (0.49)	.26	.84	.28	0.60 (0.49)	.29	.79	.40
BIF4	0.67 (0.47)	.40	.83	.45	0.69 (0.46)	.35	.79	.51
BIF5	0.57 (0.50)	.44	.83	.47	0.57 (0.50)	.40	.79	.56
BIF6	0.62 (0.49)	.42	.83	.46	0.59 (0.49)	.41	.79	.59
BIF7	0.70 (0.46)	.33	.83	.36	0.72 (0.45)	.36	.79	.54
BIF8	0.42 (0.49)	.45	.83	.49	0.46 (0.50)	.39	.79	.55
BIF9	0.64 (0.48)	.42	.83	.46	0.65 (0.48)	.39	.79	.55
BIF10	0.68 (0.47)	.29	.83	.31	0.70 (0.46)	.32	.79	.47
BIF11	0.15 (0.35)	.40	.83	.44	0.18 (0.38)	.33	.79	.57
BIF12	0.75 (0.44)	.43	.83	.47	0.77 (0.42)	.31	.79	.48
BIF13	0.58 (0.49)	.41	.83	.44	0.64 (0.48)	.35	.79	.52
BIF14	0.58 (0.49)	.39	.83	.43	0.64 (0.48)	.36	.79	.51
BIF15	0.48 (0.50)	.54	.82	.59	0.49 (0.50)	.35	.79	.51
BIF16	0.70 (0.46)	.48	.83	.52	0.77 (0.42)	.43	.79	.66
BIF17	0.36 (0.48)	.52	.83	.57	0.40 (0.49)	.46	.78	.67
BIF18	0.58 (0.49)	.29	.83	.32	0.58 (0.49)	.31	.79	.43
BIF19	0.36 (0.48)	.43	.83	.48	0.42 (0.49)	.29	.79	.42
BIF20	0.94 (0.25)	.19	.84	.21	0.93 (0.26)	.13	.80	.26
BIF21	0.49 (0.50)	.37	.83	.41	0.50 (0.50)	.28	.79	.41
BIF22	0.86 (0.35)	.18	.84	.20	0.86 (0.35)	.02	.80	.03
BIF23	0.50 (0.50)	.43	.83	.47	0.52 (0.50)	.44	.79	.62
BIF24	0.72 (0.45)	.36	.83	.39	0.74 (0.44)	.31	.79	.46
BIF25	0.82 (0.38)	.29	.83	.31	0.83 (0.37)	.26	.79	.41
Total	15.20 (5.15)	0.37^a	.84^b		15.69 (4.67)	.33^a	.80^b

BIF: Behavior Identification Form; CFA: confirmatory factor analysis; EFA: exploratory factor analysis.

^aAverage corrected item–total correlation.

^bCronbach’s alpha coefficient.

Next, EFA was used to identify meaningful factors underlying the Polish version of the BIF. EFA was performed using Sample 2 (n = 342 after listwise deletion) and evaluated using CFA in Sample 3 (n = 672 after listwise deletion). Settings for EFA (principal axis analysis) were chosen based on the original study (Vallacher & Wegner, 1989). The suitability for analysis was examined by Bartlett’s test of sphericity, χ² (300) = 1417.37, p < .001, and the Kaiser-Meyer-Olkin measure of sampling adequacy (.86). The analysis revealed the existence of one primary factor with an initial eigenvalue of 5.22, with six minor factors, with initial eigenvalues that ranged from 1.40 to 1.10. The number of factors to be retained was guided by several criteria, including Kaiser’s criterion, inspection of the scree plot, parallel analysis, and interpretability of the factor structure. The convergence of these criteria favored the one-factor structure. The finding that 23 of the 25 items had their highest loadings on the primary factor and that the two remaining items (Items 22 and 25) had positive loadings on this factor as well suggested that the BIF is indeed a unidimensional scale. Moreover, the alternative multifactor solutions that were examined showed problematic cross-loadings and were generally uninterpretable. The factor loadings from EFA with forced one-factor solution are presented in Table 3.

The factorial validity of the Polish version of the BIF scale was then tested using CFA with a weighted least-squares mean and variance (WLSMV) adjusted estimator. The results are reported in Table 3. The unidimensional model provided good fit indices, χ² = 405.75, df = 275, p < .001, comparative fit index (CFI) = .96, root mean square error of approximation (RMSEA) = .03 (90% CI: lower bound = .02, upper bound = .03), standardized root mean square residual (SRMR) = .07. All of the factor loadings, but one (Item 22), were significant at p < .001, and all, except two (Items 20 and 22), were above .40. In addition, multigroup CFA was performed to test measurement invariance of the Polish BIF with respect to gender. In line with Muthén and Asparouhov (2002), full measurement invariance was defined as all thresholds and factor loadings being the same across gender groups (i.e., scalar invariance). The baseline model, where all factor loadings and thresholds were allowed to vary freely between men and women, fitted the data reasonably well (Table 4). Model comparison showed a significant difference between the baseline model and the full measurement invariance model, Δχ²(23) = 38.22, p = .024. However, freeing Item 1’s threshold resulted in a nonsignificant difference compared with the baseline model, Δχ²(22) = 30.62, p = .104, and an acceptable model fit. These results suggest partial measurement invariance across gender.

Table 4.

Summary of goodness of fit indices and model comparisons for invariance tests.

Model	χ²(df)	CFI	RMSEA (90% CI)	Δ CFI	Δ RMSEA
Full: Baseline	670.53 (550)	.96	.03 (.02, .03)	–	–
Full: Measurement invariance	711.93 (573)	.95	.03 (.02, .03)	.006	.001
Full: Measurement invariance^BIF1	701.59 (572)	.95	.03 (.02, .03)	.003	.000
Short: Baseline	151.65 (130)	.98	.02 (.00, .04)	–	–
Short: Measurement invariance	170.22 (141)	.98	.03 (.00, .04)	.006	.003

The model marked with superscripted item number is a partial measurement invariance model. BIF: Behavior Identification Form; CFI: comparative fit index; RMSEA: root mean square error of approximation.

After examining the structure of the BIF, the performance of individual items was evaluated using internal, external, and judgmental criteria (Stanton et al., 2002) in order to develop a shortened version of the instrument.

The measures of internal item quality included average item–total correlations (across Samples 1, 2, and 3), items’ factor loadings from EFA (from Sample 2), items’ factor loadings from CFA (from Sample 3), and differences between the expected and observed counts in each category (from Sample 3). Items 1, 3, 20, 22, and 25 had the lowest corrected item–total correlations averaged across Samples 1, 2, and 3 (avg. r_it < .30). Items 3, 20, and 22 had the lowest EFA factor loadings (β < .30), and Items 20 and 22 had the lowest CFA factor loadings (β < .40). Items 5, 11, and 22 were most frequently implicated in a variable pair with large deviations between the expected and observed counts in the CFA analysis (χ² > 3.00).

The index of external item quality reflected the performance of each item relative to criterion variables conceptually related to level of action identification. The criterion variables were self-control and internal–external locus of control (Fujita, Trope et al., 2006; Vallacher & Wegner, 1989).¹ The index was calculated as an average item–criterion correlation. Items 4, 7, 10, 20, 21, 22, and 25 were found to perform poorly on this index (avg. r_pb ≤ .07).

The judgmental measures of item quality involved assessments of content validity (i.e., the extent to which each item was representative of the action identification level construct) and readability (i.e., the ease with which each item could be understood “by a person who has slightly less education than you do”). The ratings of item quality were completed by two psychologists using a seven-point scale from 1 “not representative/not very easy to understand” to 7 “very representative/very easy to understand,” and then averaged. Overall, all of the items were judged to have good content validity and readability, although Item 22 was judged as a slightly poorer indicant of the action identification level than the other items (avg. rating = 5.5), and Item 15 received slightly lower readability ratings than the other items (avg. rating = 6.5).

Based on the above findings, a 13-item Polish BIF was constructed in which the problematic items (Items 1, 3, 4, 5, 7, 10, 11, 15, 20, 21, 22, and 25) were deleted. Reliability and confirmatory factor analyses were performed on the 13-item scale and are reported in Table 5. To allow direct comparison with the 25-item measure, these analyses were performed on Sample 3.

Table 5.

Descriptive statistics, reliability information, and CFA factor loadings.

Item	M (SD)	r _it	α if item deleted	CFA
BIF2	0.88 (0.32)	.26	.72	.50
BIF6	0.59 (0.49)	.36	.71	.54
BIF8	0.46 (0.50)	.38	.70	.56
BIF9	0.65 (0.48)	.37	.70	.56
BIF12	0.77 (0.42)	.30	.71	.49
BIF13	0.64 (0.48)	.34	.71	.53
BIF14	0.64 (0.48)	.33	.71	.50
BIF16	0.77 (0.42)	.42	.70	.68
BIF17	0.40 (0.49)	.45	.69	.68
BIF18	0.58 (0.49)	.30	.71	.45
BIF19	0.42 (0.49)	.28	.72	.42
BIF23	0.52 (0.50)	.41	.70	.62
BIF24	0.74 (0.44)	.29	.71	.45
Total	8.07 (2.91)	.35^a	.72^b

BIF: Behavior Identification Form; CFA: confirmatory factor analysis.

^aAverage corrected item–total correlation.

^bCronbach’s alpha coefficient.

The Cronbach’s alpha reliability for the 13-item Polish BIF was .72 and was lower than for the full form, which seemed reasonable for a 48% reduction in the number of items. However, the value was still in the acceptable range. The item–total correlations were similar to those reported for the 25-item Polish BIF and ranged from .26 to .45 (avg. r_it = .35). The 13-item model fit the data better than the original model, χ² = 81.13, df = 65, p = .085, CFI = .99, RMSEA = .02 (90% CI: lower bound = .02, upper bound = .03), SRMR = .05. All of the factor loadings were significant at p < .001, and all were above .40. Measurement invariance was also tested across gender. The baseline model displayed a good fit to the data (Table 4). When all factor loadings and thresholds were set to be equal for men and women, the model fit did not deteriorate appreciably, Δχ²(11) = 16.40, p = .127, suggesting full measurement invariance across gender.

Discussion

The above reported analyses were conducted to further evaluate the psychometric properties of the Polish BIF and test the possibility of shortening the Polish BIF. The construct validity of the instrument was examined through the use of factor analysis and internal consistency. The results were in compliance with the original version and attested to both the unidimensionality and the internal consistency of the Polish BIF. Additional analyses examined the properties of the Polish BIF’s individual items and resulted in shortening of the length of the scale from 25 to 13 items. Overall, the psychometric properties of the shortened version were comparable to that of the full measure. Moreover, measurement invariance analyses demonstrated full invariance across gender for the short Polish BIF and partial invariance across gender for the full Polish BIF.

Correlates of the Polish BIF

Participants and procedure

The associations between levels of action identifications and other aspects of personality were examined on Sample 2 and Sample 3, so the participants and procedure were the same as those described for the factor analysis of the Polish BIF.

Measures

All participants completed the Polish version of the BIF, described earlier.

Cognitive failures were measured using the ERIS (Kowalczyk, 2013, 2016), a 50-item Polish language questionnaire that assesses the frequency of everyday slips and errors related to inattention and distraction, and not to a lack of knowledge or skills, nor to excessive task requirements. All items (e.g., “Absent-mindedly, you sweeten or salt something twice”) are scored on a five-point scale from 0 = never to 4 = very often and are summed to yield a total score.² The ERIS demonstrated high internal consistency in the present sample (Cronbach’s α = .95).

To assess self-control, the Polish SCS (Pilarska & Baumeister, 2018; Tangney, Baumeister, & Boone, 2004) was employed. It consists of 36 five-point scale items ranging from 1 = not at all to 5 = very much. These items pertain to control over thoughts, emotions, impulses, performance, and habit breaking (e.g., “I am good at resisting temptation”) and yield a single total score, with higher values indicating higher self-control. Scale reliability in the current sample was satisfactory (Cronbach’s α = .87).

Locus of control was measured with Rotter’s IE Control Scale (Rotter, 1966; adapted by Drwal, 1995). The scale is composed of 23 forced-choice items (with six additional filler items) each offering alternatives between internal- and external-control interpretations of various events (e.g., “People’s misfortunes result from the mistakes they make” vs. “Many of the unhappy things in people’s lives are partly due to bad luck”). A high score indicates an external orientation. Scale reliability in this sample was good (Cronbach’s α = .77).

Goal concordance and goal coherence were assessed using the personal striving construct (Emmons, 1999). Participant were asked to complete the PSQ (Emmons, 1999; adapted by Kadzikowska-Wrzosek, 2010) and list five current strivings, defined as “plans, objectives, and desires that we are typically trying to accomplish or attain.” Participants were then asked to rate the extent to which they pursue each striving for each of four reasons, using a 10-point scale (0 = not at all for this reason, 9 = completely because of this reason). The four reasons were, respectively, external, introjected, identified, and intrinsic. A goal concordance index was formed by subtracting averaged introjected and external ratings from those for intrinsic and identified reasons. Higher scores indicated greater goal concordance. Internal consistency was appraised as acceptable, with Cronbach’s α value of .71.

To evaluate goal coherence, participants were given a matrix adapted from the SIM (Emmons, 1999; Pilarska, 2016a). They were asked to refer to their striving list and rate the impact that being successful in one striving has on the other striving using a scale ranging from –2 to 2 (–2 = very harmful effect, 2 = very helpful effect). In order to facilitate statistical analysis of data, responses were recorded on a scale from 1 to 5, with 5 indicating a very helpful effect. A total score was obtained by averaging the ratings of the whole matrix. Higher scores indicated greater coherence among strivings. In the present sample, the SIM total score had a Cronbach’s α of .80.

Sense of identity was measured by the MQI-E, developed originally in Polish (Pilarska, 2012, 2016b). The MQI-E contains 45 items that assess six dimensions of sense of identity: sense of having inner contents, sense of uniqueness, sense of one’s own boundaries, sense of coherence, sense of continuity over time, and sense of self-worth. All items (e.g., “I feel that I was once a very different person than I am now”) are evaluated on a four-point scale (0 = strongly disagree/never, 3 = strongly agree/always), with higher scores indicating stronger sense of identity. The global score for the MQI is computed by averaging scores across six dimensions. The internal consistency reliability for the current sample was adequate (Cronbach’s α varied between subscales from .63 to .85, with a mean of .76; Cronbach’s α for all items was .91).

Self-concept clarity was assessed via the Polish version of the SCC Scale (Campbell et al., 1996; Pilarska & Suchańska, 2013). The SCC consist of 12 items focusing on the perceived consistency and temporal stability of self-beliefs (e.g., “In general, I have a clear sense of who I am and what I am”), each scored on a five-point scale from 1 “strongly disagree” to 5 “strongly agree.” Total scores are calculated by summing across all items, with higher scores indicating greater self-concept clarity. Cronbach’s α for the SCC was found to be .87.

Self-esteem was evaluated using the Polish adaptation of Rosenberg’s SES (Łaguna, Lachowicz-Tabaczek, & Dzwonkowska, 2007; Rosenberg, 1965). The SES is a 10-item scale based on a four-point Likert-type format ranging from 1 “strongly agree” to 4 “strongly disagree” (e.g., “I feel I do not have much to be proud of”). Overall scores are derived by summing across all items, with higher scores indicating greater self-esteem. The internal consistency of the SES in the present sample was deemed good (Cronbach’s α = .88).

Results

The correlational analyses were run twice to compare the results obtained with the full and shortened version of the Polish BIF. The full and short Polish BIF scores correlated highly at r = .94 (p < .001) and r = .89 (p < .001) for Sample 2 and Sample 3, respectively. No gender differences were found for either full, Z = 1.40, p = .159 and Z = 0.11, p = .914 for Sample 2 and Sample 3, respectively, or short BIF scores, Z = 1.68, p = .093 and Z = 1.30, p = .193 for Sample 2 and Sample 3, respectively. However, men tended to report a more internal locus of control, Z = 3.96, p < .001, less cognitive failures, Z = 3.47, p = .001, a higher degree of self-concept clarity, Z = 4.19, p < .001, higher self-esteem, Z = 3.04, p = .002, greater goal concordance, Z = 2.04, p = .041, and a stronger sense of identity, Z = 4.81, p < .001, than women. These results were broadly consistent with previous findings (e.g., Campbell et al., 1996; Łaguna et al., 2007).

Correlations between the full and short Polish BIF and the other scales were estimated to test whether level of action identification would predict a range of psychological outcomes (Table 6). In line with theoretical expectations as well as the results of prior studies (e.g., Fujita, Trope, et al., 2006; Vallacher & Wegner, 1989), both full and short BIF scores were positively correlated with self-control while negatively correlated with locus of control, indicating that high-level agents had better self-control and a more internal orientation. Replicating the findings of Freitas, Clark, Kim, and Levy (2009), level of agency was also positively related to concordance and coherence of goals that people set for themselves. Thus, those who held higher levels of action identification strived for goals that facilitated each other and were congruent with their values and interests. Again, these findings were observed for both full and short BIF scores. Further, as expected, level of agency was positively associated with global sense of identity. This was true when considering both full and short BIF scores and indicated that individuals high in action identification tended to experience themselves as whole, unique, and continuous beings. Finally, consistent with predictions and prior studies (Vallacher & Wegner, 1989), level of agency was found to be unrelated to self-esteem. However, comparing the present findings to those of Vallacher and Wegner (1989), two notable differences emerged such that no significant correlation was found between level of agency and cognitive failures and self-concept clarity. Following these surprising results, additional hierarchical regression analyses were used to examine the possibility that self-esteem would moderate the associations of action identification level with cognitive failures and self-concept clarity.

Table 6.

Correlations of level of action identification with various personality variables.

Measure	Sample	n	M (SD)	r		Fisher’s Z
Measure	Sample	n	M (SD)	BIF	BIF–Short	Fisher’s Z
Cognitive failures	2	343	68.80 (25.72)	–.07	–.08	.08
Locus of control	2	345	13.34 (4.25)	–.29***	–.34***	.66
Goal concordance	3	508	2.95 (2.14)	.19***	.20***	.18
Goal coherence	3	508	3.55 (0.52)	.14**	.10*	.70
Self-control	3	690	112.83 (16.75)	.23***	.25***	.36
Self-concept clarity	2	345	36.17 (8.53)	.03	.07	.44
Self-esteem	2	346	29.05 (4.74)	.08	.07	.20
Global sense of identity	3	693	1.88 (0.35)	.21***	.21***	.04

BIF: Behavior Identification Form.

*p < .05, **p < .01, ***p < .001.

These additional analyses were based on the recognition that many different self-beliefs held by individuals with high and low self-esteem influence how they act and react in certain situations. As noted by Vallacher and Wegner (1989), the optimal level of identification for effective performance depends on the action’s personal level of difficulty. Because of a higher uncertainty about personal capabilities and potentials, low self-esteem individuals may experience a wide variety of actions as more difficult than those high in self-esteem (e.g., Brockner, 1988). Thus, it is possible that undertaking everyday actions with relatively high-level identities in mind would facilitate performance for those with high self-esteem but may hinder performance of individuals with low self-esteem. Furthermore, on the basis of action identification theory, one would expect that when individuals adopt high-level identities, they become more aware of their own enduring, unobservable traits and thereby experience greater self-clarity (Vallacher & Wegner, 1989). However, it is also possible that low self-esteem individuals who exhibit preference for high-level identification would be more inclined to endorse generic self-uncertainty items because they match their overall evaluative attitude toward the self.

When predicting cognitive failures, the analysis yielded no main effect of level of personal agency, β = –.05, t = –0.98, p = .330, but there was a significant main effect of self-esteem, β = –.34, t = –6.76, p < .001. In addition, self-esteem significantly moderated the relationship between level of personal agency and cognitive failures, β = –.14, t = –2.79, p = .006, F(1, 339) = 7.76, R²Δ = .02. This interaction effect indicated that individuals who tended to identify their actions in high-level terms reported lower incidence of action slips than those inclined to control their actions with relatively low-level identities only if they also had high self-esteem. The interaction is plotted in Figure 1. For self-concept clarity, again, no main effect was found of level of personal agency, β = –.01, t = –0.16, p = .873, but there was a significant main effect of self-esteem, β = .51, t = 11.15, p < . 001. Self-esteem was also found to be a significant moderator of the relationship between level of personal agency and self-concept clarity, β = .10, t = 2.18, p = .030, F(1, 341) = 4.75, R²Δ = .01. However, neither of the simple slopes was significant at p < .05 (Figure 1). A similar pattern of results was found in the regression analyses with the short Polish BIF scores as the independent variable. The interaction effect for level of personal agency and self-esteem was significant for cognitive failures, β = –.17, t = –3.36, p = .001, F(1, 339) = 11.27, R²Δ = .03, and marginally significant for self-concept clarity, β = .08, t = 1.75, p = .081, F(1, 341) = 3.06, R²Δ = .01.

Figure 1.

Simple slopes of level of personal agency predicting cognitive failures and self-concept clarity for 1 SD below and 1 SD above the mean of self-esteem. BIF: Behavior Identification Form; SES: Self-Esteem Scale.

Discussion

Concerning criterion validity, there were expected correlations between the full and short BIF scores and most of the variables assessed—high BIF scores corresponded to higher self-control, greater perceived internal locus of control, greater self-concordance of the individual’s goals, less conflict within one’s system of goals, and stronger sense of identity. In other words, high-level personal agency was related to greater personality integration. The strengths of the associations were all small, but comparable to those reported by Vallacher and Wegner (1989). Also, there was no significant difference between correlation coefficients obtained by using full and short Polish BIF, which provided further evidence that the proposed short version could serve as a substitute for the full version.

However, contrary to expectation from the action identification theory, level of personal agency was not significantly correlated to either cognitive failures or self-concept clarity. This might perhaps be partly explained by the fact that the same instruments used by Vallacher and Wegner (1989) have not been replicated in this study. Another possible reason for the inconsistent results may be the omission of a moderating factor. The additional analyses presented above successfully used self-esteem as a moderator, but other psychological constructs may also influence the impact of a level of action identification. It should also be noted that the lack of the expected relationships of action identification level with cognitive failures and clarity of the self-concept may be due to the restricted age range of the sample studied. For example, theoretical reasoning and previous research indicate that individuals’ self-conceptions become more clear and internally consistent with age, at least up to age 60 (Diehl & Hay, 2011; Lodi-Smith & Roberts, 2010). Indeed, the participants’ self-concept clarity scores were quite low as compared to other studies (Campbell et al., 1996; Diehl & Hay, 2011; Lodi-Smith & Roberts, 2010).

Conclusions and limitations

Originally developed to measure individual differences in identification level across action domains (i.e., level of personal agency), the BIF (Vallacher & Wegner, 1989) is widely regarded as a useful measure of differences in construal and abstraction level and is frequently employed in research on various aspects of cognition and behavior (e.g., Dar & Katz, 2005; Smith & Trope, 2006). The BIF has so far been translated and adapted into several languages, including Chinese (Gu & Tse, 2018), French (Brown, 2014), Dutch (Smith & Trope, 2006), Japanese (Nakamura, Ito, Honma, Mori, & Kawaguchi, 2014), and Turkish (Yetişer, 2014), but has not yet been validated in Polish. In fact, although action identification theory was introduced over three decades ago (Wegner & Vallacher, 1986), the local literature has largely overlooked it. Surprisingly little has been published on the subject by Polish researchers (Kulesza, 2008; Trzópek, 2008), and academic books generally only provide a short description of the theory (e.g., Kofta & Doliński, 2000; Maruszewski, Doliński, Łukaszewski, & Marszał-Wiśniewska, 2008). Hence, it is hoped that this investigation would bring more attention to the notion of action identification and provoke more in-depth analyses of the principles of action identification theory. Research into this area is certainly important given the many implications level of personal agency holds for self-regulation and self-concept.

The reported analyses provide promising findings in favor of the validity and usefulness of the Polish BIF. However, a few limitations should be noted. First, the studies involved students only, thus raising concerns regarding the generalizability of the findings. Future work with other demographic and cultural groups should therefore be alert to the possibility of different findings. Second, the present study depended heavily on self-report, so shared method variance and response biases may have affected the results. Finally, it should be mentioned that the cross-sectional method prohibits firm causal inferences and conclusions. Considering that the Polish BIF demonstrated somewhat mixed results in the correlational analyses, further research is needed to provide more concrete evidence of its criterion validity and practical value. In particular, discriminant validity should be investigated by examining the relations between the Polish BIF and constructs presumed to be theoretically unrelated, such as intelligence and cognitive style (Vallacher & Wegner, 1989). For now, it is hoped that the developed Polish language versions of the BIF might urge researchers to study the effects and correlates of level of personal agency with Polish-speaking research samples.

Appendix 1.

Items of the Polish Behavior Identification Form

BIF1^a	Sporządzanie listy
	a. Organizowanie się
	b. Spisywanie rzeczy
BIF2^b	Czytanie
	a. Śledzenie linijek tekstu
	b. Zdobywanie wiedzy
BIF3^a	Wstąpienie do wojska
	a. Wspieranie obronności kraju
	b. Zaciągnięcie się
BIF4^a	Pranie ubrań
	a. Usuwanie nieprzyjemnych zapachów z ubrań
	b. Wkładanie ubrań do pralki
BIF5^a	Zrywanie jabłka
	a. Zdobywanie czegoś do jedzenia
	b. Obrywanie jabłka z gałęzi
BIF6^b	Ścinanie drzewa
	a. Używanie siekiery
	b. Zdobywanie drewna na opał
BIF7^a	Mierzenie pokoju pod wykładzinę
	a. Przygotowanie do zmiany wystroju
	b. Używanie miarki
BIF8^a,b	Sprzątanie domu
	a. Przejawianie swojej schludności
	b. Odkurzanie podłogi
BIF9^b	Malowanie pokoju
	a. Nakładanie farby pędzlem
	b. Odświeżanie pokoju
BIF10^a	Płacenie czynszu
	a. Utrzymywanie miejsca zamieszkania
	b. Zlecanie przelewu
BIF11	Dbanie o rośliny domowe
	a. Podlewanie roślin
	b. Upiększanie pokoju
BIF12^b	Zamykanie drzwi
	a. Wkładanie klucza do zamka
	b. Zabezpieczanie domu
BIF13^a,b	Głosowanie
	a. Wpływanie na wynik wyborów
	b. Zaznaczanie swojego wyboru na karcie do głosowania
BIF14^a,b	Wspinanie się na drzewo
	a. Uzyskiwanie dobrego widoku
	b. Przytrzymywanie się gałęzi
BIF15	Wypełnianie testu osobowości
	a. Odpowiadanie na pytania
	b. Pokazywanie, jakim się jest
BIF16^a,b	Mycie zębów
	a. Zapobieganie próchnicy
	b. Poruszanie szczoteczką w buzi
BIF17^b	Rozwiązywanie testu
	a. Odpowiadanie na pytania
	b. Pokazywanie swojej wiedzy
BIF18^b	Pozdrawianie kogoś
	a. Witanie się
	b. Okazywanie życzliwości
BIF19^b	Opieranie się pokusie
	a. Mówienie “nie”
	b. Wykazywanie się odwagą moralną
BIF20^a	Jedzenie
	a. Odżywanie się
	b. Żucie i przełykanie
BIF21	Uprawianie ogrodu
	a. Sianie nasion
	b. Zdobywanie świeżych warzyw
BIF22	Podróżowanie samochodem
	a. Śledzenie mapy
	b. Zwiedzanie okolicy
BIF23^a,b	Plombowanie zęba
	a. Dbanie o zęby
	b. Pójście do dentysty
BIF24^b	Rozmawianie z dzieckiem
	a. Uczenie dziecka czegoś
	b. Używanie prostych słów
BIF25	Naciskanie dzwonka u drzwi
	a. Poruszanie palcem
	b. Sprawdzanie, czy ktoś jest w domu

BIF: Behavior Identification Form.

^aReversed items.

^bItems included in the shortened version of the Polish Behavior Identification Form.

Footnotes

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