Soc Just Res (2011) 24:297–313
DOI 10.1007/s11211-011-0142-7
Are Perceptions of Organizational Justice Universal?
An Exploration of Measurement Invariance Across
Thirteen Cultures
Ronald Fischer • Maria Cristina Ferreira • Ding-Yu Jiang •
Bor-Shiuan Cheng • Mustapha M. Achoui • Corbin C. Wong •
Gulfidan Baris • Socorro Mendoza • Nathalie van Meurs •
Donna Achmadi • Arif Hassan • Gunes Zeytinoglu • Figen Dalyan
Charles Harb • Dania D. Darwish • Eveline M. Assmar
•
Published online: 1 December 2011
Ó Springer Science+Business Media, LLC 2011
Abstract Previous research examined whether justice effects are comparable,
focusing on quantitative differences in justice effects. This study examines whether
justice perceptions are structured similarly or whether they are qualitatively different across working populations from 13 nations. Confirmatory factor analysis and
multi-group analysis show that Colquitt’s (J Appl Psychol 86:386–400, 2001) fourdimensional model of justice works well across these samples. However, factor
intercorrelations and reliabilities are found to systematically vary between cultural
R. Fischer (&) S. Mendoza D. Achmadi
Victoria University of Wellington, Wellington, New Zealand
e-mail: ronald.fischer@vuw.ac.nz
M. C. Ferreira
Salgado de Oliveira University, Rio de Janeiro, Brazil
D.-Y. Jiang
National Chung-Cheng University, Minhsiung, Chiayi County, Taiwan
B.-S. Cheng
National Taiwan University, Taipei, Taiwan
M. M. Achoui
Arab Open University, Kuwait, Kuwait
C. C. Wong
Deutsche Bank AG, New York, USA
G. Baris G. Zeytinoglu F. Dalyan
Anadolu University, Eskisehir, Turkey
N. van Meurs
Middlesex University Business School, London, UK
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samples. Perceptions of justice are more highly intercorrelated in power distant and
collectivistic samples, in line with extensions of the relational model of authority.
Score reliabilities were lower in collectivistic settings.
Keywords Organizational justice Power distance Collectivism Equivalence
Invariance Multi-group factor analysis
Organizational justice research over the last four decades has highlighted the
importance of perceptions of justice for work behaviour and motivation in
Western societies (Cohen-Charash & Spector, 2001; Colquitt, Conlon, Wesson,
Porter, & Ng, 2001). However, the extent to which these findings can be
generalized to non-Western work contexts is still under-explored. Justice research
provides important insights into work motivation of employees (Erez & Earley,
1993); therefore, it is important to study justice perceptions in more diverse
samples to gain a better understanding of employee concerns in our globalized
world. Expatriate managers making decisions may assume that employees in nonWestern contexts may react in a similar way to their decisions as would
employees in their home country; yet, the growing cross-cultural literature has
demonstrated that these assumptions are often erroneous (see Gelfand, Erez, &
Aycan, 2007; Tsui, Nifadkar, & Ou, 2007). Given the very limited work on
perceptions of justice in a wider sample of nations, we present a first examination
of the factor structure of a widely used US instrument (Colquitt, 2001), using
multi-group confirmatory factor analysis and means-covariance structure analysis
(Cheung & Rensvold, 2000; Vandenberg & Lance, 2000) in samples from 13
diverse nations. We also examine whether cultural dimensions from previous
research (House, Hanges, Javidan, Dorfman, & Gupta, 2004; Schwartz, 2006) can
help us to explain how employees perceive justice events. With a sample of 13
countries, such an analysis by necessity will remain somewhat exploratory and
preliminary. However, since we are the first to explore these issues, we believe the
findings to be reported merit further empirical attention and by linking cultural
dimensions to psychometric indicators; we provide a novel approach to the study
of structural invariance in a cross-cultural context.
A. Hassan
International Islamic University Malaysia, Kuala Lumpur, Malaysia
C. Harb D. D. Darwish
American University of Beirut, Beirut, Lebanon
E. M. Assmar
Salgado de Oliveira University, Niteroi, Brazil
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Dimensionality of Justice in the US and International Samples
There is now relative consensus in the Western literature that four major dimensions
of justice perceptions can be distinguished (Colquitt & Shaw, 2005). The first and
oldest dimension is distributive justice which focuses on the distribution of rewards
and whether people believe that their outcomes and rewards match their input or
investments (Adams, 1965). The second component is procedural justice, which
refers to the procedures that determine these outcomes (Thibaut & Walker, 1975).
Bies and Moag (1986) introduced the concept of interactional justice. Interactional
justice is focused on the treatment of individuals by decision-makers and whether
they show respect, sensitivity and explain decisions thoroughly. Subsequently,
Greenberg (1993) separated this interactional component into the interactional
aspect of procedures, namely the adequacy of providing information (informational
justice) and the interpersonal treatment by supervisors when explaining distributive
decisions (interpersonal justice).
Colquitt (2001) developed and validated a measure differentiating these four
components. There is some evidence now that the four different components of
justice relate differently to various outcome measures in the US samples (e.g., Bell,
Wiechmann, & Ryan, 2006, Chiaburu & Marinova, 2006, Colquitt, 2001; Colquitt
et al., 2001; Judge & Colquitt, 2004; Roch & Shanock, 2006).
Very limited research in non-Western societies has been conducted. Why is it
important to examine the factor structure of justice dimensions across different
cultures? First, justice perceptions provide an important insight into the motivational states of employees (Colquitt et al., 2001; Erez & Earley, 1993). Many
employment surveys today routinely assess general perceptions of justice in
Western companies and feedback derived from these assessments influences
management decisions. The increasing globalization of the workplace means that
managers have to lead teams consisting of employees from many different cultures.
Similarly, greater world-wide standardization of HR practices means that subsidiaries in other parts of the world now also apply these employment surveys. If the
responses are structured differently in non-Western contexts or for migrants from
non-Western countries, then management interventions may waste time, money and
resources or even damage employee–management relations (see Erez & Earley,
1993).
Second, recent studies have started to explore differences in justice effects across
cultures. There is some evidence that justice effects are not uniform across cultural
contexts (Brockner et al., 2001; Farh, Earley, & Lin, 1997; Fischer & Smith, 2006;
Lam, Schaubroeck, & Aryee, 2002; for reviews, see Fischer, 2008; Gelfand et al.,
2007; Tsui et al., 2007). If there was a different factor structure or items may be
comparatively easier or difficult (meaning differences in item intercepts across
cultures), then no cross-cultural comparisons can be attempted and previous findings
are meaningless (Fontaine, 2005; van de Vijver & Leung, 1997).
Finally and extending the previous point, current research has not addressed the
universality of justice perceptions directly. Henrich, Heine, and Norenzayan (2010)
pointed out that the majority of research is conducted using Western, educated,
industrialized, rich and democratic (WEIRD) samples. These samples come from a
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limited geographical range and account for about one-quarter of the world
population. Therefore, our insights into the psychology of justice are very limited.
More research on non-Western samples is clearly needed. Therefore, we first test
whether the four-dimensional structure can be fitted to all samples.
The available data to date provides a mixed picture. To the best of our
knowledge, no study exists that has explored all four dimensions included in the
Colquitt (2001) questionnaire. Some studies with data from Taiwan (Farh et al.,
1997), Hong Kong (Fields, Pang, & Chiu, 2000) and Portugal (Rego, 2002)
supported the distinctiveness of distributive, procedural and broad interactional
justice components, whereas other studies with samples from France (Igalens &
Roussel, 1999), Britain (Fischer & Smith, 2006), Germany (Fischer & Smith, 2006;
Pillai, Williams, & Tan, 2001), Hong Kong and India (Pillai et al., 2001) did not.
As a working hypothesis we could propose that employees in all societies can
make a basic distinction between these four dimensions. These dimensions refer to
distinctions between specific outcomes and procedures that lead to these outcomes
as well as social versus structural components of these outcomes and procedures
(see Greenberg, 1993). Therefore, justice may be a functional universal (Norenzayan & Heine, 2005, see also Henrich et al., 2010). However, we do not know
whether the individual justice elements such as the procedural justice criteria
discussed by Leventhal (1980) or the Bies and Moag’s (1986) elements of
interactional justice are also universal. There is some research suggesting that these
criteria may work differently across cultures (Fischer, 2008; Leung & Tong, 2004).
Therefore, the individual items that capture these criteria in Colquitt’s scale may not
work equally well in all samples. Therefore, we are the first to test the factor
structure using multi-group confirmatory analysis across a large number of samples
(research question 1).
Exploring Differences in Justice Dimensions Across Cultures
Societies differ along a number of cultural dimensions, of which the extent to which
people are individualistic or collectivistic (individualism–collectivism: Hofstede,
2001; autonomy vs. embeddedness: Schwartz, 2006) and the extent to which is
power is expected to be distributed equally or not (power distance: Hofstede, 2001;
egalitarianism vs. hierarchy: Schwartz, 2006) appear particularly important for
justice research (see Fischer & Smith, 2003; Lind, Tyler, & Huo, 1997). The
relational model of authority (Tyler & Lind, 1992) proposes that individuals care
about procedural justice because it provides them with information about inclusion
in valued groups (being considered a fully fledged member of one’s work group).
Lind et al. (1997) were the first to show that in hierarchical societies, individuals
pay less attention to procedural justice information. This is because their relative
status within the group is determined by the cultural context and therefore the
context overrides the need to evaluate justice information to evaluate their standing
within the group. To the extent that concerns about hierarchical standing is
important, power distance may be an important moderator. In contrast, if the
primary is concern with inclusion in the group overall, individualism–collectivism
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301
would also be important. Subsequent research then demonstrated that this context
effect is stable (Brockner et al., 2001; Farh et al., 1997; Fischer & Smith, 2006; Lam
et al., 2002; for a review, see Fischer, 2008) and also applies to other justice
dimensions, including distributive justice (Fischer & Smith, 2004).
These studies have shown that justice perceptions may not be as informative for
individuals in some contexts compared to others. Although this does not imply that
individuals will not be able to discriminate between the four dimensions, it may
have some implications for the factor structure, specifically the factor correlations.
The informative value of justice is decreased and, therefore, the motivation of
individuals to make such fine-grained differentiations may be reduced (as implied
by models of humans as goal oriented information processing agents, De Dreu,
Nijstad, & van Knippenberg, 1998; Forgas & George, 2001). Drawing upon these
information processing models, it could be expected that justice dimensions may
appear more interchangeable and similar in more hierarchical and collectivistic
contexts. Therefore, we propose that latent justice factors are more highly
intercorrelated in more power distant, hierarchical and collectivistic settings
(hypothesis 1).
The Current Study
We are testing the cross-cultural applicability of the four-dimensional scale
developed by Colquitt (2001) across samples from 13 different countries. The
samples were selected in an effort to maximize variability along the dimensions of
power distance/hierarchy and collectivism/embeddedness as measured in two recent
multi-national projects (House et al., 2004, Schwartz, 1994, 2006). Schwartz (1994,
2006) measured values in teacher samples from around the world. We have samples
covering the whole range of his collectivism equivalent dimension of embeddedness
versus autonomy dimension. House et al. (2004) distinguished between practices
and values when measuring institutional and ingroup collectivism as well as power
distance. Our samples capture the whole range of variability along the ingroup
collectivism practices and values and the power distance values. Our samples are
somewhat higher in power distance samples because these contexts have been
relatively ignored in previous justice research.
Method
Samples
Participants were recruited through personal contacts of the authors, local
collaborators and snow-ball sampling. In each location, attempts were made to
maximize variability in terms of occupation, organizational status, industry and
sector. Surveys were distributed to contact persons within organizations (who often
made additional copies) and completed surveys were returned to the contact person
or directly to the local investigator via pre-paid envelopes. Surveys were always
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Table 1 Demographic information on study samples
Effective N
Mean age (SD)
%Male
%Managers
Argentina
215
39.11 (10.28)
47.0
18.6
13.0
Brazil
275
35.92 (9.52)
50.5
32.7
33.2
Egypt
75
31.08 (8.59)
44.6
55.2
.0
196
34.98 (9.76)
61.5
56.0
25.2
Indonesia
%Public sector
Lebanon
398
33.21 (9.68)
52.0
53.6
9.1
Malaysia
284
33.55 (7.74)
41.7
70.8
54.2
51.5
New Zealand
141
37.91 (10.59)
52.4
37.2
Philippines
628
35.82 (9.52)
53.5
63.0
5.1
Saudi Arabia
239
32.11 (7.87)
85.7
29.5
10.8
Taiwan
365
32.27 (7.93)
55.2
32.5
69.9
Turkey
148
31.81 (8.40)
53.8
40.4
23.5
United Kingdom
161
45.18 (9.72)
32.8
10.0
100.0
United States of America
145
34.01 (12.19)
48.4
29.2
55.1
completed outside work hours. It was stressed that completion and submission of the
questionnaire was entirely voluntary and that answers were treated anonymously.
The data collection method did not allow an exact calculation of response rates.
In total, 3,283 completed surveys with no missing information were returned.
Data was available from samples in Argentina, Brazil, Egypt, Indonesia, Lebanon,
Malaysia, New Zealand, Philippines, Saudi Arabia, Taiwan, Turkey, UK, and US.
The sample sizes ranged between 75 in Egypt and 628 in the Philippines. The
average sample size was 251.5. Table 1 reports more information about the
demographic characteristics of the samples.
Instruments
Organizational Justice
We used the organizational justice measure developed by Colquitt (2001). This
scale measures procedural, distributive, interpersonal and informational aspects of
organizational justice. It has been developed and validated in the US (Colquitt,
2001; Judge & Colquitt, 2004). Answers were recorded on seven-point likert scales
with the labels ‘‘(1) Not at all’’, ‘‘(4) To some extent’’ and ‘‘(7) To a great extent’’.
Participants were asked to respond to questions regarding some decision made in
their workplace. Seven questions focused on procedural justice which was presented
first. An example question is ‘‘Have those procedures been free of bias?’’
Interpersonal justice was measured with four items. The questions were directly
related to supervisor/superior actions and an example question is ‘‘Has s/he treated
you with dignity?’’ Informational justice focuses on the perceived adequacy of
explanations given to respondents by their supervisor/superior and was measured
with five items. An example question is ‘‘Has s/he explained the procedures leading
to a decision thoroughly?’’ Distributive justice was presented last. Four items
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Table 2 Sample score
reliabilities (Cronbach’s alpha)
PJ procedural justice, DJ
distributive justice, IPJ
interpersonal justice, IFJ
informational justice
PJ
DJ
IPJ
IFJ
Average alpha
Argentina
.71
.89
.86
.91
.84
Brazil
.82
.92
.86
.89
.87
Egypt
.87
.95
.94
.92
.92
Indonesia
.86
.91
.88
.87
.88
Lebanon
.90
.94
.92
.94
.92
Malaysia
.88
.90
.92
.92
.90
New Zealand
.86
.89
.92
.90
.89
Philippines
.90
.94
.91
.91
.91
Saudi Arabia
.87
.89
.85
.94
.89
Taiwan
.89
.96
.95
.95
.94
Turkey
.88
.92
.89
.91
.90
UK
.90
.92
.96
.94
.93
US
.88
.91
.94
.92
.91
measuring perceptions of the outcomes from decisions that are made in the
organization were included. An example question is ‘‘Does the outcome of these
decisions reflect the effort you have put into your work?’’
The English version of the questionnaire (with adaptations to the local language
and writing) was used in the US, UK, New Zealand as well as India, Malaysia and
the Philippines. The English version was used in these contexts since English is an
official language in these countries and is widely spoken in the business community.
Arabic and English versions were simultaneously used in Lebanon, Egypt and Saudi
Arabia. The English version was translated into Arabic (Saudi Arabia, Egypt,
Lebanon), Portuguese (Brazil), Spanish (Argentina), Bahasa Indonesia (Indonesia),
Mandarin (Taiwan) and Turkish (Turkey) using a series of translation–backtranslation and committee translation procedures. This combined method is superior to
simple translation–backtranslation (Van de Vijver & Hambleton, 1996).
Table 2 displays the reliabilities per sample. As can be seen there, the overall
reliabilities are all acceptable and beyond the commonly accepted threshold of .70
(Nunnally, 1978). In fact, the overall reliability across all samples and dimensions is
.90, suggesting very good internal consistencies.
Country Level Indicators
We used scores for societal level cultural indicators measuring power distance/
hierarchy and collectivism/embeddedness from both the GLOBE project (House
et al., 2004) and Schwartz (1994, 2006). House et al. (2004) distinguish between
ingroup collectivism (familism), institutional collectivism and power distance
societal practices and values. We therefore tested effects for both societal practices
and values. Data for all nations is available, except for Lebanon and Saudi Arabia.
We therefore used averages for other Arabic samples as estimates for these two
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countries. We repeated all analyses excluding these two countries and the patterns
were similar.
Data Analysis
We tested whether the four-factor structure fitted in the individual samples fitted in
comparison to a standard solution (see Fischer & Fontaine, 2010 for further
discussions of options). We chose the US sample as the standard solution (since the
questionnaire was developed in the US) and ran a multi-group CFA. We used
maximum-likelihood estimation in LISREL 8.52. To evaluate fit, we used a number
of fit indices in addition to the Chi-squared statistic which has been shown to have
problematic properties (e.g., Bentler, 1990, Bentler & Bonnett, 1980; Bollen, 1989;
Mulaik et al., 1989). First, the comparative fit index (CFI, Bentler, 1990) and the
Tucker-Lewis Fit index (TLI) or non-normed-fit-index (Bentler & Bonnett, 1980)
were used. The TLI has been found to be very robust and relatively sample size
independent (Mulaik et al., 1989). Values ranging above .95 have traditionally been
seen as indicating an acceptable fit (Hu and Bentler, 1999; Marsh, Balla, &
McDonald, 1988). A second set of indices are lack of fit indices. The Root Mean
Square Error of Approximation (RMSEA; Browne & Cudeck, 1992) is considered
because it punishes less parsimonious models. A value of less than .05 is ideal,
values ranging between .06 and .08 are acceptable and values larger than .10
indicate poor fit. Finally, the Standardized Root Mean Square Residual (SRMR) is
used since it has been shown to be the most sensitive fit index detecting simple
model misspecifications (Hu & Bentler, 1999). Values smaller than .08 are
indicating approximate fit.
In line with traditional equivalence/invariance approaches in organizational and
cross-cultural research (e.g., Fontaine, 2005; van de Vijver & Leung, 1997;
Vandenberg & Lance, 2000), we compare increasingly restricted models. This
analysis answers whether the instrument can be used in cross-cultural research and
whether scores can be directly compared. The first level tested is configural or
structural equivalence (indicating weak factorial invariance, Meredith, 1993). The
loading patterns are constrained to be similar across groups. The next step is metric
equivalence or measurement unit equivalence, tested by constraining the factor
loadings to be identical across cultural groups. The next level is full score or scalar
equivalence. Item intercepts are constrained to be identical across groups. Once this
level of equivalence has been shown, scores can normally be directly compared
across cultural groups. These levels of equivalence are the most important ones to
be tested in cross-cultural research (van de Vijver & Leung, 1997; Vandenberg &
Lance, 2000).
We used deterioration in model fit between the original four-dimensional model
of justice and the hierarchically constrained models as indicators of misfit.
Following Cheung and Rensvold (2000) and Little (1997), we used DTLI equal or
less than .01 as indicating similarly fitting models.
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Further Exploration of Cultural Differences
Additional levels of equivalence/invariance can be tested and are of theoretical
interest for our purposes. First, invariance of uniqueness is interesting because it
relates to reliability estimates of scores. It implies that the unique variance
(measurement error) is identical across cultures. In case this assumption is rejected,
the scales show different reliabilities across samples. This may be of some
importance for practitioners as increased unreliability reduces validity and therefore
may explain why justice is less strongly correlated with work variables in some
contexts. We will further explore reliability differences (see below).
Finally, invariance of the latent variable variance/covariance matrix can be
tested. In this case, the variances and covariances of the latent factors are
constrained to be equal across groups. If these constraints do not lead to a
substantive deterioration of model fit, intercorrelations between the latent factors are
similar. In this case, individuals in all our samples do perceive justice in the same
manner, independent of cultural background (Fontaine, 2005). If we have to reject
equality of factor intercorrelations, we can proceed and test hypothesis 1. As above,
we use the relative fit of each model and reject a model if the deterioration in model
fit for DCFI/TLI is larger than .01.
To test our hypothesis 1, we used correlations. First, we averaged the latent
correlations across four justice dimensions (see Tables 3, 4). We then correlated this
average intercorrelation with the culture-level dimensions. If we found a significant
correlation with the averaged factor correlation, we then explored the nature of the
correlation by examining the correlations between pairs of latent justice factor
correlations and the culture-level dimensions. We chose this strategy as there is a
relatively large number of correlations that need to be run and our sample of
Table 3 Latent factor intercorrelations
PJ 9 IPJ
PJ 9 IFJ
IPJ 9 IFJ
PJ 9 DJ
IPJ 9 DJ
IFJ 9 DJ
Average
correlation
Argentina
.66
.73
.77
.36
.21
.11
.47
Brazil
.61
.61
.78
.57
.52
.56
.61
Egypt
.47
.64
.79
.54
.42
.64
.58
Indonesia
.54
.70
.65
.77
.53
.71
.65
Lebanon
.53
.63
.76
.71
.52
.71
.64
Malaysia
.63
.71
.77
.66
.53
.73
.67
New Zealand
.63
.68
.81
.71
.49
.61
.66
Philippines
.58
.66
.83
.73
.63
.73
.69
Saudi Arabia
.52
.61
.69
.62
.57
.66
.61
Taiwan
.64
.67
.81
.62
.56
.58
.65
Turkey
.73
.79
.83
.83
.68
.83
.78
UK
.49
.64
.72
.87
.42
.52
.61
US
.56
.66
.86
.68
.55
.67
.66
PJ procedural justice, DJ distributive justice, IPJ interpersonal justice, IFJ informational justice
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Table 4 Correlations of
average reliabilities and factor
intercorrelations with countrylevel indicators
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Average
alpha
Hierarchy (vs. egalitarianism)
.52*
Autonomy (vs. embeddedness)
-.17
.13
Power distance practices
-.34
-.01
Power distance values
.48
.07
Institutional collectivism practices
.34
.47
Ingroup collectivism practices
* p \ .05
.22
Average factor
intercorrelations
.03
Institutional collectivism values
-.50*
Ingroup collectivism values
-.23
.23
-.34
.50*
countries is relatively small, making family-wise error rates a threat to our analysis.
It should be noted that this strategy is more conservative (because it may miss
dimension-specific effects that are obscured in the average factor intercorrelation).
We used Spearman rank-order correlations due to the non-normality of many
country-level indicators and interpret correlations exceeding |.50|, indicating a large
effect size.
Results
Testing the Four-Factor Structure (Research Question 1)
We first tested whether the four-dimensional structure fits in all 13 samples equally
well. The multi-group analyses tested whether factor patterns, factor loadings, and
factor intercepts are invariant or equivalent across cultural groups.
Using the US as reference group, we first tested whether factor loading patterns
were similar across groups. This multi-group model fitted reasonably well: v2
(2132) = 6238.56, RMSEA = .088, TLI = .97, CFI = .97. All items in all
samples loaded significantly on the expected factor. On average, each sample
contributed about 7.69% to the overall v2, except in the case of the Philippines
(15.63). SRMR in each sample was below .08. In the case of the Philippines, the
contribution to the v2 was relatively large, but the SRMR was the lowest.
Constraining the factor loadings to be identical next, the fit was acceptable: v2
(2324) = 7060.53, RMSEA = .09, TLI = .97, CFI = .97. The DTLI and DCFI
were both 0, indicating no substantive decrease in fit between models. Only the
RMSEA was at the higher end of acceptable fit. Again, the Philippines contributed
more to the overall v2 than other samples (14.74%).
In the final step, we constrained item intercepts to be equal. The fit was still
acceptable: v2 (2564) = 7814.93, RMSEA = .09, TLI = .97, CFI = .96. The
deterioration of fit was only minimal (DTLI = 0, DCFI = .01). Again, the Filipino
sample contributed relatively more to overall sample fit than other samples
(13.83%). Levels of SRMR were comparable to the previous model (model B). The
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net difference between the two models is less than .001 suggesting that constraining
the intercepts did not strongly impact on residuals. Overall, this suggests that factor
loadings and variables intercepts are sufficiently similar. This level of scalar or full
score equivalence allows comparison of means across cultural groups. To examine
the relative contribution of the Filipino sample to the model misfit, we correlated
sample size with fit indicators for each sample. Sample size correlated strongly with
v2 (q = .89, p \ .001) and SRMR (q = -.58, p \ .05). This confirms previous
findings of sample size dependence of v2 and explains why the large Filipino sample
was contributing relatively disproportional to the overall v2. In summary, overall
there is strong evidence that the four factors fit well across the 13 cultural samples
included in our study.
Additional Invariance Tests
We tested two more models, one in which we constrained the unique variances to be
invariant and a second model in which latent factor variances and covariances were
constrained. First, the model testing for invariance in measurement error at the item
level (unique variances) fitted considerably worse than the previous models: v2
(2804) = 17296.36, RMSEA = .14, TLI = .92, CFI = .91. Therefore, the fit was
worse than for the previous model constraining only the intercepts (DTLI = .05,
DCFI = .06). Overall, this suggests that reliabilities are different across cultural
groups since measurement error is not invariant. We will explore these differences
below.
The final model tested latent variable variance–covariance matrix equivalence.
The fit of this model was somewhat worse than the fit for the model testing full
score equivalence: v2 (2684) = 9633.43, RMSEA = .102, TLI = .96, CFI = .96.
Although the differences for TLI and CFI were small (.01), RMSEA deteriorated
above the threshold of .10. Overall, this indicates that correlations between the
latent variables are not strictly equivalent. Since the lower levels of invariance were
generally acceptable, this difference cannot simply be explained by measurement
artefacts, but suggests some genuine differences in psychological processes. We can
therefore proceed with testing our hypothesis 1.
Exploring Differences in Factor Intercorrelations (Hypothesis 1)
Correlating the average correlation across all four justice dimensions with countrylevel indicators, the correlation with ingroup collectivism values (q = .50, p = .08)
and hierarchy (versus egalitarianism) values (q = .52, p = .06) had moderate effect
sizes. Greater collectivism and hierarchy values were associated with stronger
correlations, i.e., is greater similarity in justice perceptions. This supports
hypothesis 1. This effect was particularly strong for the procedural justice–
informational justice link and ingroup collectivism values (q = .69, p \ .05), the
procedural justice–distributive justice link and hierarchy values (q = .61, p \ .05),
and interpersonal justice–distributive justice links and hierarchy values (q = .61,
p \ .05). Overall, the pattern of results shows support for hypothesis 1.
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Exploring Differences in Score Reliabilities
We followed a similar procedure as described for the intercorrelations. We first
examined the correlations between the Cronbach alpha coefficients (average of the
four alphas per dimension and sample, see Table 2) and the culture-level
dimensions. Two effects were significant. The averaged alphas were correlated
with institutional collectivism values (q = -.50, p = .07). Greater institutional
collectivism values were associated with lower reliabilities for procedural justice
(q = -.55, p = .05) and interpersonal justice (q = -.58, p \ .05).
Discussion
Our study contributes to the literature in three ways. First, we provide a test of a
widely used US scale across a wide range of cultural samples. We are the first to test
these dimensions comprehensively in a larger number of non-Western samples,
covering samples from all inhabited continents and all major cultural and religious
groups in the world. The most important finding of our study is that this measure of
justice is applicable across these cultural samples. The good fit for the fourdimensional structure provides broader support for claims that these dimensions are
indeed empirically distinct. Employees around the world seem to make the same
distinctions.
Second, we found that latent factor score variances and covariances differ across
cultures. Our exploration of latent factor intercorrelations shows that in hierarchical
and collectivistic settings justice dimensions are somewhat more strongly
correlated. This provides support for our speculation that although individuals in
all societies have the capacity to differentiate between the four dimensions (see the
good fit of the four-dimensional structure in the multi-group analysis), there is less
informative value contained in these justice perceptions for individuals in
hierarchical and employees may therefore have a somewhat lower motivation to
distinguish these latent variables. This fits our extensions of the relational model of
authority (Tyler & Lind, 1992) to the perception of justice dimensions. This
somewhat greater overlap in justice perceptions may have some practical
implications. For example, behaviour of organizational decision-makers and
managers is evaluated more holistically by employees in these contexts compared
to more egalitarian and individualistic settings, where employees are motivated to
make finer differentiations between individual justice components.
Third, we also found that score reliabilities are high, but not identical across
cultural groups. The societal context has a significant and important influence on the
measurement process. Schmitt and Allik (2005) reported that self-esteem measures
showed higher reliability in more individualistic and lower reliability in more power
distant settings. Here, we found somewhat similar effects. It is unclear what drives
these effects. One reviewer provided an interesting hypothesis. Individuals in
collectivistic societies are more concerned with fitting in. Therefore, they should be
motivated to avoid extreme answers, which leads to restricted variance in more
collectivistic and hierarchical societies (Johnson, Kulesa, Cho, & Shavitt, 2005;
123
Soc Just Res (2011) 24:297–313
309
Smith, 2004, 2011; Smith & Fischer, 2008). Variance restriction is one factor that
can influence internal consistency measures such as Cronbach’s alpha and
correlations more generally. In line with this argument, we also found a significant
correlation between the overall variance of the justice scores and in-group
collectivism values (q = -.64, p \ .05). In more collectivistic societies, the
variance is more restricted. However, the same restriction of variance argument
would imply that intercorrelations between factors are likely to be weaker in
collectivistic societies. We found the exact opposite. As a consequence, it is unclear
what accounts for these differences. Motivations to fit in collectivistic settings
leading to restricted variance may be one factor, but this is inconsistent with other
aspects of the data. In short, our study replicates previously noted effects (Schmitt &
Allik, 2005), but we cannot offer a plausible explanation for the overall patterns in
our data (increased factor correlations but decreased reliability in collectivistic
settings). Clearly, we need more research on these effects, especially considering
the implications for human resource initiatives because differential score reliabilities will influence confidence intervals and validities (Bollen, 1989). Differential
effects of justice dimensions on variables (e.g., in regressions) may be due to
differential reliabilities rather than substantive effects. We offered one option for
investigating these effects and future work needs to extend our analyses.
Concerning data analysis, the statistical techniques used here are sophisticated
but are now widely available in programmes such as LISREL, AMOS, MPlus and
EQS. Application of these techniques in cross-cultural settings in conjunction with
further data exploration (either using correlations as in our case or multi-level
modelling) can be used to address important theoretical questions (see for example
Lucas et al., 2008). We strongly recommend greater adoption of these techniques in
cross-cultural justice work.
Limitations
We relied on fit indices when evaluating model fit. Judging misfit in multi-group
analyses is ambiguous (Cheung & Rensvold, 2000; Vandenberg & Lance, 2000).
We used deterioration in the fit indices as indicator of misfit, however, no
significance tests (exact tests) are available and researchers have to rely on
heuristics such as the DTLI equal or less than .01 criterion (Cheung & Rensvold,
2000). Second, it is possible that other items could be relevant for measuring justice
across cultures. Issues of domain underrepresentation (Fontaine, 2005) cannot be
ruled out through statistical testing. This requires qualitative enquiries in each
cultural context to examine whether additional items are important for measuring
each dimension of justice in the specific context. A third limitation is that the items
are reasonably abstract. It is not clear whether the enactment of procedures such as
providing appeal mechanisms or treating employees with dignity and respect entails
the same behaviours across cultures (Leung & Tong, 2004; see Fischer, 2008 for a
review). To date, there is little information on cultural enactment of these justice
principles. This is a serious gap in the literature. Fourth, other principles within each
dimension are plausible. The dimension that is best understood in a cross-cultural
123
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Soc Just Res (2011) 24:297–313
context is distributive justice. The scale we used was based on the US research that
only included an equity component. However, other principles such equality or need
might be important and salient in different cultures. Fischer et al. (2007) developed
such a distributive measure incorporating equality and need in addition to equity
and reported significant correlations of need and equality, but not equity with both
societal values and economic indicators. Future scale development exercise should
pay attention to include alternative principles of relevance in non-Western contexts.
Finally, the sampling was based on convenience. Therefore, the samples were not
directly matched across cultures. However, we believe this is not an issue. The
countries in our study are at different stages of economic and industrial
development. Restricting the sampling of participants to specific industries would
not capture the diversity of economic activities within each of the countries. For our
hypothesis we proposed specific cultural dimensions and we have selected samples
so that they represent the whole spectrum of the relevant dimensions as best as
possible. Therefore, we were able to test whether these cultural dimensions explain
factor intercorrelations and score reliabilities across samples. Previous research has
demonstrated that non-matching (e.g., due to industry, occupational and sample
differences) is likely to attenuate cultural differences (Fischer & Chalmers, 2008;
Fischer & Mansell, 2009; Schmitt, Allik, McCrae, & Benet-Martinez, 2007). Since
we found significant effects of cultural dimensions despite the non-matching of
industries and occupations, we can be relatively confident in the strength of the
cultural effects.
Nevertheless, our study is an important step forward in justice research. We
demonstrated that a widely used scale is applicable in different cultural contexts.
We provide an example of testing the structure and applicability of the US scales in
non-Western settings. The scale shows full score equivalence, allowing researchers
to compare scores directly between countries. This is a major achievement and
facilitates future work on cross-cultural differences in justice perceptions. However,
we also found some systematic differences in factor intercorrelations. These
findings point to some consistent, albeit weaker influence of culture on the
experience and expression of justice.
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
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The Practice of Social Research, 14e
PRINTED BY: alexbonilla813@gmail.com. Printing is for personal, private use only. No part of this book may be reproduced or transmitted without
publisher's prior permission. Violators will be prosecuted.
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11/9/2018
The Practice of Social Research, 14e
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