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Determinants and Effects of Implicit and Explicit Comparison Standards on Salary Demands: Gender Differences and Beyond
Rez Shirazi, Anders Biel, and Niklas Fransson Department of Psychology Göteborg University, Sweden
Shirazi, R, Biel, A, and Fransson, N., Determinants and Effects of Implicit and Explicit Comparison Standards on Salary Demands: Gender Differences and Beyond , Göteborg Psychological Reports, 2002, 32, No. 4. This research examined effects of perceived common fate, similarity, proximity, and availability of comparison targets and comparers’ motives on the choice of social comparative referents. Moreover, the relationship between social and internal comparison standards and demanded salaries for similar others was assessed. Data results from 904 employees indicated that internal comparison standards as allocation anchors completely mediated gender differences in pay allocations. Implications for the gender wage gap and managers’ salary decisions in general are discussed. Key words: Allocation anchor, gender differences, salary demands, comparison standards
In past decades behavioral scientists have shown a burgeoning interest in the study of structural or individual determinants of salary differentials (e.g., gender salary differentials: Roos & Gatta, 1999). At the individual level, researchers have studied antecedents of the factual salary gap such as education and career path, preferences and expectancies of work outcomes, particularly salary expectations and current demands. A major rationale for studies of salary demands is the long-lasting and accumulative effects of the individual and collective salary demands, especially at the initial stages of entry into the labor market, on actual salary and other monetary and non-monetary fringe benefits (e.g., health insurance, retirement plans, sickness and unemployment compensations) (e.g., Gerhart & Rynes, 1991). ___________________________________________________________________________ Author Note: This study is part of a research program supported by the Swedish Council for Social Research’s Grant 98-0242 to Anders Biel. We thank Riël Vermunt, Tommy Gärling, and Lisbeth Hedelin for their comments on an early version of this paper, and Susanna Nisser for their assistance in conducting the study. We also wish to thank those organizations and individuals who kindly volunteered to participate in the study. Correspondence concerning this article should be addressed to Rez Shirazi, Department of Psychology, Göteborg University, Box 500, SE 405 30 Göteborg, Sweden. Electronic mail may be sent via the
No. 4:32, 2 Internet to
[email protected]. Researchers in this area have paid special attention to the study of gender differences in pay expectations and demands, partly due to the fact that a gender wage gap remains (e.g., Roos & Gatta, 1999) despite decades of legal, social, and organizational policies to narrow the gap. The interest in the study of gender differences in pay expectations and demands also stems from the act that several studies have indicated that women expect lower pay than their male counterparts (e.g., Major & Konar, 1984) and report high satisfaction with their earnings in spite of objective criteria pointing to their disadvantage (e.g., Crosby, 1982). This phenomenon is sometimes referred to as the “depressed-entitlement effect” among women (Jost, 1997) or, more frequently, as the “paradox of the contented female worker” (Crosby, 1982: 12). However, the empirical support for the occurrence of gender differences in pay allocations and related evaluations (e.g., pay satisfaction and fairness evaluations) and proposed mediators has been mixed and equivocal. One influential line of explanation has emphasized gender differences in biased social (intra-group and inter-personal) comparisons as mediators of gender differences in pay allocations and related evaluations (e.g., Crosby, 1982; Jackson, Gardner, Sullivan, 1992; Major, 1989; Major & Konar, 1984; McFarlin, Frone, Major, & Konar, 1989). It is suggested that because women, compared with men, make more intra-group comparisons of salaries (i.e., choose other women as wage-comparative referents) and women generally have lower salaries than men have, women report lower pay expectations and demands. However, as we argue in the following, gender differences in pay allocations in previous studies may be explained predominantly by differences in internal comparison standards. This explanation suggests a similar and parsimonious process for both genders and can be generalized beyond gender differences to other group and individual differences in pay expectations and demands. Thus, the current paper presents and examines a general explanation of differences in salary allocations with an emphasis on the mediating effects of implicit and explicit internal comparison standards. The proposed explanation is used to incorporate certain conflicting results of previous research on gender differences in pay allocations as a specific example of its applicability. Moreover, we replicate and extend prior research on the choice of comparative referents. We also explore the strength and validity of the concept of entitativity (“the degree of having the nature of an entity”), originally proposed by Campbell (1958: 17), in predicting the choice of comparative referents.
Theory and hypotheses A number of studies, predominantly using student samples, indicate that women, relative to men, report lower pay expectations and demands (Bylsma & Major, 1992, 1994; Jackson et al., 1992; Jost, 1997; Keaveny & Inderrieden, 2000; Major & Konar, 1984; Major, McFarlin, & Gagnon, 1984; Martin, 1989; McFarlin et al., 1989). Research also shows that pay expectations affect subsequent pay allocations (Major, Vanderslice, & McFarlin, 1984). Major and
No. 4:32, 3 associates (Major, McFarlin, & Gagnon, 1984, Exp. 2) demonstrated that when the researcher paid a fixed amount of money for a certain job prior to the start of a job, women worked significantly longer and performed better than did men. However, several studies have reported unexpected, conflicting, or null results with regard to gender differences in intra-group comparisons and pay allocations (e.g., Gasser, Flint, Tan, 2000; Major & Testa, 1989, Exp. 2; Major, Vanderslice, & McFarlin, 1984). Although several leading social and pay evaluation theories (e.g., Adams, 1963; Crosby, 1976; Festinger, 1954; Lawler, 1971) refer to the influence of the individual’s internal (e.g., personal past) or external (e.g., other person or group) comparison standards on perceived inequity, self-evaluations, and feelings of deprivation or satisfaction , the fundamental question regarding the selection of comparative referents remains unanswered (see also Greenberg, 2001). More specifically, certain critical issues that researchers have partly addressed pertain to the effects of group membership on comparison processes (Ambrose & Kulik, 1988), the type of occupational outcome (e.g., pay) or procedure being compared (Ambrose, Harland, & Kulik, 1991), and personal characteristics (e.g., tenure) of the comparer (Oldham, Kulik, Stepina, & Ambrose, 1986; Oldham, Nottenburg, Kassner, Ferris, Fedor, & Masters, 1982). There are comprehensive reviews of determinants of the choice of comparison standards and referent typologies in a wide range of contexts (e.g., Levine & Moreland, 1987) and particularly in organizational settings (e.g., Kulik & Ambrose, 1992). Although there are several common elements in approaches and results of empirical studies, attempts to develop theories of the choice of comparative referents in work settings have been few (e.g., Gartrell, 1982; Shah, 1998). In the following, we propose a cognitive approach to integrate past research and generate predictions. This approach shares two concepts (proximity, similarity) with other perspectives and is meant to complement previous approaches. In the past decade, social psychologists have increasingly utilized the construct of entitativity (Campbell, 1958) in research on inter- and intra-group cognition and behavior. One line of research is concerned with determinants and processes that lead to perceptions of an aggregate of individuals as a social entity or a group, and the other is concerned with effects of the perceived entitativity. Applied to the research on comparison standards, we suggest that the choice of comparative referents is determined by the degree to which a comparer perceives himself/herself and a potential comparative referent as being members of the same entity (e.g., occupational, organizational, and social entity). Campbell (1958) suggested four sources of entitativity ranked by their influence, namely, common fate with the greatest influence on the perceived entitativity followed by similarity, proximity, and lack of permeability. We examine effects of the first three on the choice of comparative referents. In work settings, occupational commonality as an instantiation of common fate functions as a strong source of perceived group entitativity and drawing group boundaries within and between organizations. For instance, Moore’s (1991) study showed that respondents most frequently mentioned job incumbents within their occupations, followed by employees in their organizations, as the referents for comparison of their salaries. Furthermore, various dimensions of similarity draw group boundaries and divide employees
No. 4:32, 4 into different entities. Due to occupational segregation by gender within and between organizations (e.g., Anker, 1997; Jacobs, 1999; Roos & Gatta, 1999) and general social roles and stereotypes, gender is still a strong source of perceived similarity within groups, and boundary between groups, in work settings. Perceived entitativity results in preference for in-group members (e.g., samegender co-workers) rather than out-group members (e.g., other-gender coworkers) for wage comparison purposes. Thus, we crossed the occupational commonality defining common fate with gender similarity and built four combinations with the same-occupation and same-gender combination defining the highest group entitativity, due mainly to the predominance of occupational commonality. Empirical evidence supports this predicted order. Major and Forcey (1985; see also Major & Testa, 1989) found that participants ranked information about same-job others for comparison of wages with highest preference. More importantly, participants showed stronger preference for wage information about same-job other-gender others than same-gender other-job others. Notably, men and women did not differ in their social comparison preferences. The strong effect of the preference for in-group members on the choice of comparative referents may be moderated by the availability of in-group versus out-group individuals (cf. visibility in Weick, 1966; and proximity, visibility, and accessibility in Major, 1989). Under conditions in which out-group individuals are more numerous and, therefore, more available than in-group members, the preference for in-group comparative referents may be less pronounced and employees may more frequently choose a comparative referent of the opposite gender. Specifically, women working in male-dominated occupations (i.e., 70% or more male employees) may more frequently choose male co-workers as comparative referents than do women working in female-dominated occupations. Similarly, male respondents’ choice of other men, relative to women, as comparative referent, would be more sizable in male-dominated occupations and less sizable in female-dominated occupations in which there are relatively few men available for comparison. Hypothesis 1. In work settings, employees’ choice of comparative referents is primarily determined by occupational commonality followed by gender similarity, other factors being equal. Thus, we predict that (a) respondents will more frequently choose individuals within their occupational and gender group as comparison targets than individuals outside their occupational and/or gender group; (b) Due to stronger effect of occupational commonality than gender similarity on the perceived entitativity in work settings, employees more frequently choose a comparative referent of the opposite gender within their occupation than a same-gender referent in other occupations; (c) Frequency of cross-gender comparisons increases as members of the other gender group become more numerous and, therefore, more available for comparison. The extent to which an aggregate of individuals are physically closer to each other affects the degree of perceived entitativity of that aggregate as a group. When occupational commonality or lack thereof is constant among a number of employees, physical proximity of a comparison target to the comparer determines the choice of the comparison target. There is limited research on the effect of physical proximity on comparison processes, and results have been
No. 4:32, 5 mixed (e.g., Shah, 1998). We examine the main effect of proximity and its interaction with occupational commonality. Thus, Hypothesis 2. In work settings, employees’ choice of comparative referents is determined by physical proximity, other factors being equal. Hence, employees will more frequently choose employees in their workplace than employees outside their workplace as comparative referents. Motives for and direction of social comparisons. In addition to perceived group entitativity, motives for comparisons may also determine the selection and direction of comparisons (e.g., Levine & Moreland, 1987; Wood, 1989). In a review of the research on social comparisons of personal attributes, Wood (1989) suggested that self-enhancement or self-protective goals motivate people to use comparative referents that are inferior to them (downward comparisons) in order to cope with threatening situations or personal disadvantages and shortcomings (see also Levine & Moreland, 1987). In contrast, comparisons with superior others (upward comparisons) may result in selfimprovement through setting higher standards and aspirations for selfevaluation. In the context of equity evaluations, Levine and Moreland (1987) stated that self-improvement motives and resultant upward comparisons are likely when comparers perceive that they can take advantage of relative inferiority in negotiations for higher outcomes (e.g., salary). Although upward comparisons may be uncomforting and aversive, Wood concluded that most studies indicate preferences for lateral or upward comparative referents rather than downward comparison (Wood, 1989, see also Oldham et al., 1982). Results of Moore’s (1991) study showed that respondents spontaneously selected comparative referents who they believed earned higher salaries than the respondents did. Furthermore, in a study of determinants of CEO compensation, O’Reilly, Main, and Crystal (1988) reported a sizable influence of the highest salary among the members of compensation committee on CEO cash compensation. Thus, we predict that motives to improve one’s own conditions precede self-protective goals and result in a preference for upward comparisons. Moreover, due to the hypothesized predominance of common fate (e.g., occupational commonality) on the choice of comparative referents, we predict that employees more frequently select a comparative referent within their occupation, though inferior on the comparison dimension (e.g., salary), rather than a comparative referent who is superior on the comparison dimension but different in occupational tasks (see the discussion on two interpretations of similarity in social comparison research in Wood, 1989: 234-236). Finally, with regard to effects of the direction of comparisons, we predict that upward comparisons raise aspiration levels and subsequent pay allocations to self and similar others whereas downward comparisons result in lowered outcome standards and ensuing pay allocations. Hypothesis 3. (a) Within occupation, employees more frequently choose others with highest salaries as their comparative referents than those with lowest salaries; (b) Employees also select lowest-paid employees in their occupation more frequently than those with higher salaries but in different occupations (e.g., their managers). (c) Upward comparisons, comparison with highest-paid within occupation or managers, increase respondents’ salary demands for others in their
No. 4:32, 6 occupation while downward comparisons, comparisons with lowest paid employees in the occupation, decrease their salary demands for others. Allocation anchors as implicit comparison standards. In the early 70s, Tversky and Kahneman’s influential research (e.g., 1974) on decision heuristics and biases introduced the anchoring and adjustment heuristic. Both field and laboratory studies have shown the effect of anchoring on pay allocations and evaluations. O’Reilly and associates (1988) applied the findings from studies on the anchoring effect and social comparisons in a study of CEO compensation when the influence of standard economic determinants was accounted for. Of direct relevance to the present study, their findings supported the hypothesis that individuals on the board of directors and the compensation committee are selected on the basis of their similarity to CEOs, and they may anchor their compensation decisions for a CEO in, and by comparing to, their own salaries. Results showed that salaries of the members of the compensation committee were the best predictor of CEO compensation. In a follow-up article (Belliveau, O’Reilly, & Wade, 1996), when the researchers added measures of social similarity and status to analyses, a measure of social comparison (the average salary of the outside executives on the compensation committee) remained a significant predictor of CEO compensation. In controlled settings of experiments, Markovsky (1988) demonstrated that participants’ pay allocations to others (and fairness evaluations) could be manipulated in a predicted direction and biased toward certain values by providing participants with initial values or anchors (e.g., minimum wage, highest wage, or wage of a similar other). In a frequently cited article, Major and Konar (1984) introduced and tested a five-factor model of gender differences in pay expectations, finding strong support for the influence of internal comparison standards on management students’ expected salary at career entry and career peak. Men, compared to women, perceived higher salaries for others in their field and also expected higher salaries for themselves. More importantly, gender differences in pay expectations were diminished when the influence of comparison standards was accounted for. Similarly, Major, McFarlin, and Gagnon (1984, see also Bylsma & Major, 1992, 1994; Major, Vanderslice, & McFarlin, 1984) found that gender differences in self-pay behavior diminished when information about others’ pay was made available (for an exception, see Martin, 1989). Furthermore, at variance with the predicted same-gender choice of pay information, the amount of self-pay was equal to the average amount taken by other (ostensible) participants regardless of their gender. If provision of starting comparison standards, or anchors, affects ensuing pay allocations, the question then becomes: What happens when participants are not given any initial anchors? Extrapolating from the results of studies on CEO compensation (Belliveau et al., 1996; O’Reilly et al., 1988) to low-paid and/or blue-collar employees, we suggest that when the allocator and the target share contextually relevant attributes (e.g., occupation and gender), the allocator primarily relies on internal comparison standards and anchors (e.g., his/her own current earnings) for pay allocations, particularly in the absence of explicit external anchors. The resultant pay may then be adjusted upward or downward according to additional factors such as group and individual motives (e.g., selfprotective versus self-improvement motives). Hence, the question here is: When
No. 4:32, 7 gender gap in actual salaries and resultant women’s lower initial anchors is taken into account, do women adjust their pay allocations, such as salary demands, upward and sufficiently to bridge the gap? The answer to this question is of pivotal importance because it may reflect the degree of awareness of the gender wage gap between and within occupations, or of being underpaid among low-paid and/or blue-collar employees. Such awareness is a precondition for employees’ reasonable demands and bargaining for increased salaries for these occupations relative to other occupations, as well as for female employees relative to male co-workers within occupations. In summary, the results of the studies reviewed above lend indirect support to the proposition that gender differences in pay allocations may be explained by the differing internal comparison standards of men and women and that these differences diminish when allocation anchors are either experimentally (cf., Desmarais & Curtis, 1997a; Major, McFarlin, & Gagnon, 1984) or, as in the present study, statistically accounted for (cf., Desmarais & Curtis, 1997b). Nonetheless, on the basis of the mixed results of previous research on gender differences in pay allocations, and the fact that we intend to study a working sample in low-paid and/or blue-collar occupations in an occupational and societal context with one of the highest ratios of women’s to men’s earnings in the world (90% in Sweden relative to 71% in the US, Roos & Gatta, 1999: 104), we do not pose any hypothesis about the occurrence of gender differences in pay allocations. However, we predict that whenever there is a gender difference in pay allocations, employees’ internal comparison standards mediate such a gender difference. Thus, Hypothesis 4. Any gender difference in pay allocations is mediated by differences in internal comparison standards. Employees’ own current or past salaries function as internal anchors for pay allocations, which are then adjusted upward or downward.
Methods Setting, Sample, and Procedures Three organizations, two labor unions and a large hospital, agreed to participate in a study on “job evaluation and salary decision making”. The first organization was the Swedish Metalworkers’ Union with 21% female membership (Statistics Sweden, 1998: 201). Next, the Commercial Employee’s Union organizes employees in the wholesale and retail (hereinafter: W&R) trade industries (71% women: Statistics Sweden, 1998: 201). The third organization was northern Europe’s largest hospital with approximately 17,000 employees (83% women), of which 5,800 were nurses/equivalent and 4,700 auxiliary nurses/equivalent. The three organizations provided lists of members/employees who had been sampled randomly from membership or employment directories. With regard to the hospital subsample, we had required only lists of nurses and auxiliary nurses because of our interest in low-paid occupations. Thus, 1,739 union members or hospital employees received a questionnaire (52.3% returned). Data analyses were based on the final sample of 904 respondents, 333 men (Age:x = 40.7, s.d. = 10.55) and 571 women (Age:x = 43.7, s.d. = 9.69).
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Questionnaire On the cover page of the booklet, we informed participants about the aim and content of the study and the research ethics we comply with (e.g., confidentiality of responses). Participants did not receive compensation for their participation. The questionnaire consisted of two mixed parts. One was constructed to address a different theoretical issue and is thus omitted here. Items in the second part of the booklet addressed determinants of differences in pay allocations, including gender differences, and mediation of comparison standards.
Measures Independent variables: Gender and subsample membership. Respondents reported their gender, which was dummy-coded (Man = 0, Woman = 1). The gender composition of the occupation constituted the second independent variable (male-dominated [70% or more male employees] vs. female-dominated occupations). This categorization was based on the consensus and organizational data reported above. The gender ratio of the occupation has direct implications for Hypothesis 1c, which predicted an increase of cross-gender comparisons as the relative number of members of the opposite gender, and thus their availability, increases. The gender ratio of the occupation also has implications for Hypothesis 4 predicting a relationship between participants’ current and previous salaries and their mediating effects as internal comparison standards, and pay allocations. As evident in prior research (e.g., Anker, 1997; Jacobs, 1999), male-dominated occupations are generally better paid than femaledominated occupations. Moreover, women in male-dominated occupations have higher salaries than their counterparts in female-dominated occupations. Due to assumed group variability differences between the W&R Trade and Hospital subsamples and observed differences in preliminary analyses, we decided to use the subsample membership as the independent variable. Hence, the Metall subsample represents male-dominated workplaces (79% men) and the W&R Trade (71% women) and Hospital (83% women) subsamples represent femaledominated workplaces. Control variables. We statistically controlled for differences in two measures of human capital endowments. Participants reported years of job experience in their current occupations and their highest level of education, ranging from elementary school (1) to university degree (7). Mediators. Mediators consisted of explicit (i.e., reported by the respondent) and implicit comparison standards. The verbatim instructions for explicit comparison standards were the following: “To arrive at what constitutes a reasonable salary for one’s own work many [employees] compare what they earn and what other [employees] earn. Some [employees] compare their salaries with employees within the same occupation or same workplace; some compare their salaries with individuals of the same gender, etc. In the following, we have listed a number of groups with which one may compare one’s own salary. For each one, indicate whether you usually compare your salary with that group or
No. 4:32, 9 not.” The instructions were followed by a list of 14 potential comparison targets (e.g., “men in other occupations”), of which 13 were forced-choice items. Each of these 13 items (12 social and one internal target) was followed by three response alternatives (Yes, No, and Don’t know/Not sure). Responses were dummy-coded and used in analyses in which active choices were coded as positive responses (Yes = 1) and the remaining as negative (No/Don’t know/Not sure/Missing response = 0). Item 14, an open-ended question, gave the respondent the opportunity to indicate other potential comparison targets not included in the list. However, few respondents used this option and their responses were thus omitted. Social comparison standards. Twelve forced-choice items measured social comparison standards. Of these 12 items, nine pertained to the three sources of entitativity and the remaining three items corresponded to crossing occupational commonality and the direction of comparisons (upward vs. downward). Common fate and similarity were operationally defined by two dichotomous variables, occupational commonality (same-occupation vs. otheroccupation) and gender similarity (same-gender vs. other-gender) between the respondent and a potential comparative referent. These two variables were crossed and built four combinations (see the upper portion of Table 3). Moreover, physical proximity between the respondent and the referent was defined by the proximity of workplaces (same-workplace vs. other-workplace). This variable was crossed with occupational commonality forming a four-cell combination (see the lower portion of Table 3, excluding the last row). Because Swedes have the opportunity to work in the neighboring Scandinavian countries, we included a fifth cell, representing a potential comparative referent working in the sameoccupation abroad. Hence, nine items measured the relationship between the three sources of the perceived group entitativity and the choice of comparative referents, which were used as predictors of pay allocations. Three items measured the relation between occupational commonality, the direction of selected comparisons and the choice of comparative referents: Comparisons of one’s own salary with salaries of (a) lowest-paid employees, (b) highest-paid employees within the comparers’ occupation, and (c) their managers, respectively. The downward equivalent to the last item (e.g., comparison with one’s subordinates’ salaries) was not applicable to the intended sample because few employees had supervisory positions. Internal comparison standards. Participants reported their own current monthly salary before taxes. This variable was used as the measure of implicit internal standard. Participants also indicated whether they usually compared their salary to what they had earned previously. This forced-choice item was dummy-coded (Yes = 1, No/Don’t know/Not sure/Missing response = 0) and used as the measure of explicit internal comparison standard. Dependent variables. We instructed participants to indicate the amount of monthly salary employees in their occupation should receive when entering the occupation and after five years in the occupation, respectively. We used the measure of salary after five years in the occupation to reduce unknown variance in pay allocations by controlling for the length of organizational tenure and onthe-job training. A comparison between predictors of demanded starting salary and salary after five years in the occupation facilitates the examination of
No. 4:32, 10 generalizability of a regression model predicting pay allocations that differ in the time frame. Analyses We used nonparametric statistical analyses of frequencies of positive (yes) responses to explicit comparison items to test the hypotheses about the relationship between three sources of group entitativity and the choice of comparative referents (Hypotheses 1a-3b). We conducted hierarchical multiple regression analyses to (a) check for the occurrence of the depressed entitlement effect among women in the current sample, and (b) test Hypotheses 3c and 4, predicting the effects of social and internal comparison standards on salary demands for others and the predicted mediation of any gender difference in pay allocations.
Results Table 1 provides descriptive statistics for all variables. Table 2 breaks out the means and standard deviations for the variables by subsample and gender group. Preliminary analyses revealed that members of the Metalworkers Labor Union, particularly men, had the highest monthly salary, whereas female members of the W&R Trade Union reported the lowest current salary. With regard to the explicit comparison standards, 86.2% of participants made multiple choices (n = 779,x = 6.27, s.d. = 3.62) as measured by the number of positive responses to the 13 explicit comparison items (12 social and one internal), whereas 9.4% of participants indicated no explicit comparison and 4.4% indicated one comparison only. As presented in Table 2, participants in the Hospital subsample reported making more explicit comparisons than did participants in the two remaining subsamples.
Table 1 a Means, Standard Deviations, and Correlations for Men and Women Mean s.d. 1 2 Variable 15.16 2.43 1. Demanded starting salaryb 15.18 2.53 18.81 2.98 .79** 2. Demanded salary after five b 18.72 3.08 .81** years
3
4
16.66 2.81 .33** .41** 15.66 3.04 .52** .46** 4. Education 4.90 1.96 .28** .21** .08 5.13 2.02 .36** .39** .33** 5. Job experience 14.51 10.71 -.08 -.07 .15** -.37** 16.80 9.57 .04 -.04 .16** -.18** aFor each variable, means, standard deviations, and correlations for men (n = 333) are shown on the first line; comparable statistics for women (n = 571) are presented on the second line. 3. Current salaryb
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measures represent amounts in thousands Swedish crowns (SEK 10 = 1
USD). ** p < .01. Hypothesis 1a stated that employees more frequently select employees within their occupation than those in other occupations as comparative referents. Regardless of the gender and workplace of the referent, the sum of positive responses to the four items regarding comparative referents within the comparer’s occupation (see the left portion of Table 3, excluding the last row) was 2,235 (61.8% of 3,616 possible positive responses to four items), while the comparable figure for the four items regarding referents working in other occupations than the comparer’s was 1,476 (40.8%). There was strong support for Hypothesis 1a. A McNemar test for paired-samples revealed that respondents more frequently selected employees within their occupation than those in other occupations for salary comparisons, regardless of the gender similarity or workplace proximity; ? 2 (N = 3,616) = 429.74, p < .001.
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Hypothesis 1a also predicted that the choice of same-gender individuals as comparative referents would significantly exceed that of the opposite-gender. Regardless of the referent’s occupation, a McNemar test indicated that respondents more frequently selected a referent of similar gender (47.5%) than one of the opposite gender (37.0%); ? 2 (N = 1,808) = 120.68, p < .001. To examine gender differences in the choice of comparative referents and women’s biased intra-group comparisons suggested by some scholars, we partitioned the positive responses to same-gender within-occupation referent (n = 503, 55.6% of 904 possible positive responses to one item) according to the respondent’s gender. The chi-square test for two independent samples did not indicate that women more frequently chose other women in their occupation (57.4%, n = 328) than men selected other men (52.6%, n = 175) for wage comparison purposes; ? 2 (N = 904) = 2.04, ns. To test Hypothesis 1b, which stated that occupational commonality exerts stronger influence on the choice of the comparative referent than does gender similarity, we conducted a pairwise contrast with a Bonferroni correction. A McNemar test showed that, contrasting the frequency of the choice of sameoccupation, other-gender comparative referents (45.6%, n = 412) with the frequency for other-occupation, same-gender comparative referents (39.4%, n = 356), respondents more frequently chose individuals with whom they shared an occupation rather than a gender group (? 2 [N = 904] = 9.28, p < .01). Thus, as predicted, the results indicated that shared occupation and gender determined the choice of comparative referents, with a predominance of the effect of shared occupation. Hypothesis 1c stated that the availability of comparative referents moderates respondents’ preference for in-group (e.g., same-gender) comparisons. As shown in Table 4, disregarding the gender composition of the occupations, male respondents’ choice of comparative referents decreased significantly when the potential referent target changed from a male to a female referent (the McNemar test for the significance of changes: ? 2 [N = 333] = 40.98, p < .001). Partitioning this overall effect according to the subsample membership, post-hoc analyses showed that, as expected, male respondents’ same-gender preference was sizable and significant in the male-dominated occupations (the Metalworkers’ Union), whereas men working in the female-dominated workplaces (the Hospital) did not differentiate between male and female referents in their occupation. Unexpectedly, men in the female-dominated W&R Trade subsample reported sizable preference for same-gender referents.
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No. 4:32, 15 Hypothesis 1c also predicted a similar pattern for female respondents’ preference for same-gender referents. Regardless of the gender composition of employees in the occupations, female respondents’ selection of comparative referents increased significantly when the referent target changed from a male to a female referent (the McNemar test: ? 2 [N = 571] = 17.28, p < .001). As presented in the lower portion of Table 4, we partitioned female respondents’ overall preference for same-gender referents to separate analyses for the three subsamples. As predicted, female respondents working in female-dominated occupations at the Hospital or in the W&R trade sector significantly preferred a female referent in their occupation while women in male-dominated occupations (the Metalworkers’ Union) did not indicate a sizable preference for same-gender comparative referents. In sum, there was overall support for the prediction that the preference for in-group members as comparative referents is moderated by the availability of in-group versus out-group comparative referents. There was partial support for the prediction that physical proximity is related to the choice of a comparative referent (Hypothesis 2). Results of the Cochran Q test for related samples indicated that the three sets of frequencies of the choice of the comparative referent with varying physical proximity to the respondent (within the occupation in the same workplace, in other workplaces, and abroad) differed significantly among themselves; Q (df = 2)= 450.96, p < .001. In post-hoc analyses, the McNemar test with a Bonferroni correction for three pairwise comparisons indicated that comparative referents abroad were less frequently selected than referents within (? 2 [N = 904] = 252.78, p < .001) and outside (? 2 [N = 904] = 286.69, p < .001) the comparer’s own workplace, respectively. Unexpectedly, there was no sizable difference between the choice of comparative referents working in the same and other workplaces. Hypothesis 3a stated that, within occupations, employees more frequently compare their own salary with that of employees with highest salaries than those with lowest salaries. Furthermore, it was predicted (Hypothesis 3b) that for salary comparisons, employees more frequently select employees in their occupation with lowest salaries than their own managers who work in dissimilar occupations but are better paid. To test these predictions, we compared positive responses to the item measuring comparison of one’s own salary relative to those of employees with highest salaries (55.8%, n = 504), lowest salaries (36.3%, n = 328), and to managers’ salaries (28.5%, n = 258), respectively. Results of the Cochran Q test for related samples indicated that the three sets of frequencies differed significantly among themselves; Q (df = 2)= 222.61, p < .001. Pairwise comparisons using the McNemar test with a Bonferroni correction for three posthoc tests indicated that all three pairs differed significantly and in the predicted order (ps < .005). To examine gender differences in upward and downward comparisons, we compared women’s and men’s positive responses to the three comparison items. Three separate chi-square tests for two independent samples did not indicate any sizeable gender difference in the comparisons with highestpaid employees (? 2 [N = 904] = 2.06, ns), lowest-paid employees (? 2 [N = 904] = 0.55, ns), or managers (? 2 [N = 904] = 0.22, ns). Hypothesis 4 predicted that any gender difference in pay allocations is mediated by differences in comparison standards, particularly by differences in internal comparison standards. Moreover, it was predicted that employees’
No. 4:32, 16 current salaries function as implicit comparison standards or anchors for their salary demands for similar others. To test this hypothesis, we performed two hierarchical regression analyses of demanded starting salary and demanded salary after five years in the occupation, respectively (see Table 5).
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No. 4:32, 18 As shown in Table 5, controlling for the relationships between participants’ education and job experience and their salary demands for others, results showed a significant main effect of gender on both dependent variables. Women demanded lower starting salary and salary after five years for others in their occupation. Furthermore, employees in the W&R trade occupations demanded significantly lower starting salaries than the average for the three subsamples while employees at the Hospital reported significantly higher salary demands for others. The mean starting salary demands for employees in the male-dominated occupations (the Metalworkers’ Union) did not differ from the average for the three subsamples. (The Subsample membership factor was represented by two independent variables using an effects coding method: the Metalworkers subsample was assigned the string of -1s, the W&R Trade Union = 1 and the Hospital = 0 in the first IV, and 0 and 1 in the second IV, respectively). However, the main effect of gender on demanded starting salary was modified by a significant interaction in Step 3. The observed gender difference was mainly due to women’s lower demanded starting salary in the W&R Trade subsample in comparison with their male counterparts. With regard to the salary demands for others after five years, the interaction between gender and subsample predictors was not significant, indicating that women in all three subsamples, relative to men, uniformly and similarly demanded lower salaries for others after five years in the occupation. As shown in the left portion of Table 5, adding a set of 12 variables representing social comparisons to the regression model increased the explained variance significantly, though by only 3%. The gender gap in starting salary demands remained significant (b = -425.33, SE b= 172.07, β = -.08, p < .05). Of all respondents, those who reported that they usually compared their salaries with same-gender employees in their occupation demanded lower starting salaries for others. In support of Hypothesis 3c, respondents who compared their own salary with lowest-paid employees in their occupation (downward comparisons) reported lower starting salary demands for similar others. Moreover, a positive relationship was evident for comparison of one’s own salary with managers’ salaries (upward and dissimilar comparisons) and demanded starting salaries. Unexpectedly, there was no sizable relationship between comparisons with highest-paid employees (upward and similar comparisons) and demanded staring salary. Therefore, Hypothesis 3c was partially supported. Hypothesis 4 which predicted that internal comparison standards mediate any gender difference in pay allocations was fully supported. Adding two explicit and implicit indicators of internal comparative referents in Step 5 increased the explained variance significantly and by 11%. Moreover, the significant gender difference observed in the previous steps diminished (β = .00, b = 25.90, SE= 164.05, ns). Results revealed that participants who usually compared their salary with their previous salaries reported lower starting salary demands for others. There were 554 positive responses to this item (61.3% of 904 possible positive responses), which constituted 9.8% of all observed positive responses to the 13 explicit comparisons (N = 5,670). To examine possible gender differences in internal comparisons, we used the chi-square test for two independent samples. There was no sizeable difference between women’s (60.4%, n = 345) and men’s (62.8%, n = 209) choice of their previous salaries as comparison standard;
No. 4:32, 19 ? 2 (N = 904) = 0.49, ns. As predicted, respondents’ current salary as implicit allocation anchors exerted strong and unique influence on demanded starting salary for others (sr = .33, β = .31, p < .001). In fact, the two internal comparison standards, first and foremost respondent’s current salary, not only mediated the effect of gender on demanded starting salary but also mediated the relationship between several social comparison standards and demanded starting salary. The latter mediation was indicated by the decrease of the relationships presented in Table 5 (Step 4) with the minor exception of the positive influence of comparisons with salaries of employees abroad on salary demands, which remained significant at the 5% level. As shown in the right portion of Table 5, results of the regression of our second major dependent variable, demanded salary for others after five years in the occupation, on the same predictors were generally in parallel to the results reported for demanded starting salary. However, some results are of importance and deserve comment. First, results of upward and downward comparisons as predictors of the second dependent variable were stronger and even upward comparison of one’s own salary with highest-paid in the occupation emerged as a significant predictor of demanded salary for others after five years in the occupation. Hence, Hypothesis 3c was fully supported. Second, not only did results reveal that respondents’ current salary functioned as implicit allocation anchors for pay allocation to similar others, there were also significant indications that participants, especially women, adjusted their allocations upward. The upward adjustment was evident when we matched a subgroup among respondents (n = 319) with the average of job experience comparable to that instructed in our second measure of pay allocation (salary after five years) and compared their actual salary (x = 15,576 SEK, s.d. = 3,040) with their salary demands for their counterparts after five years in the occupation (x = 18,886 SEK, s.d. = 3,218). These respondents had an average of 5.6 years of job experience (range = 9.8, s.d. = 3.18). A 2 (gender) X 2 (respondent’s current salary vs. demanded salary after five years) analysis of variance with repeated measure on the latter factor revealed that participants significantly increased their salary demands for others relative to what they actually earned (an adjustment upward); F1, 317 = 357.05, p < .001. More importantly, as indicated by a significant interaction, the increase of salary for others demanded by female participants (n = 169) was significantly more than that demanded by male participants; F1, 317 = 10.67, p < .01. In fact, women indicated a salary increase to such an extent that if both groups’ salary demands had been met the actual gender wage gap (men:x = 16,267 SEK; women:x = 14,962 SEK) would have diminished (men:x = 18,978 SEK; women:x = 18,805 SEK). The pattern of the results was similar across the three subsamples, but a stronger increase in demanded salary was observed for the employees at the Hospital. Disregarding the matching criterion of five years’ job experience, the pattern of an upward adjustment of the salary demands for others after five years in the occupation was similar for the three subsamples and for the entire sample. Robustness of results. To establish robustness of the results, we added two-way interaction terms to the regression equations in Step 6. The interaction terms included gender by social comparison standards and gender by internal
No. 4:32, 20 comparison standards. Neither the ∆R2 (the incremental variance explained) for the starting salary nor for the demanded salary after five years was significant (F change for both measures: F14, 868 < 1). Thus, there was no indication that men’s and women’s choices of social comparison standards, particularly intragroup comparisons, were differentially related to pay allocations.
Discussion The most significant result of this study is support for our hypothesis that internal comparison standards mediate differences in pay allocations and the employee’s own pay functions as implicit allocation anchor in their salary demands. Equally interesting was the finding that employees, particularly women, adjusted these salary demands upward (cf. the assimilation effect, Markovsky, 1988) and sufficiently to bridge existing pay differentials. Although the importance of current pay on pay expectation and satisfaction has been noted previously, many researchers have neglected to include current salary level in their studies of pay expectations and pertinent evaluations, partly due to a lack of a theoretical explanation and/or interest. As noted elsewhere (Keaveny & Inderrieden, 2000; Sweeney, McFarlin, & Inderrieden, 1990), the omission of such a strong and influential predictor may have produced distorted estimates of other predictors. Even when researchers had included participants’ current pay in their studies and established its dominant relationship with pay allocations and evaluations, they have considered those findings “not theoretically interesting”. We believe that the concept of anchoring and adjustment provides a compelling framework to understand conflicting findings of past research and to explain present and future results. The consistency among the results of field studies on CEO compensation (e.g., Belliveau et al., 1996; O’Reilly et al., 1988), laboratory experiments on college students’ pay allocations (Markovsky, 1988), and this study, in which the compensation of low-paid and/or blue-collar occupations was examined, lends strong support to the generalizability of the relationship between allocation anchors as implicit comparison standards and diverse pay allocations and evaluations. The relationship is strongest when the allocator and the allocation target are similar or peers, such as members of a compensation committee and a CEO (e.g., Belliveau et al., 1996; O’Reilly et al., 1988) or, as in the present study, experienced employees and new colleagues in the occupation. The Case of Gender Differences in Pay Allocations To demonstrate and test the explanatory value of internal comparison standards as referential anchors in explaining group differences in pay allocations, the depressed entitlement effect among women represented a welldocumented group difference of both theoretical and practical importance. Our results replicated the occurrence of gender difference and indicated that women, relative to men, reported lower salary demands for others in their occupations. More importantly and in support of our hypothesis, the observed gender differences diminished first and only after partialling out the differences in internal standards, mostly in current salaries. In contrast, the alternative
No. 4:32, 21 explanation that argues for the biased intra-group and/or intra-personal comparisons did not receive support. First, men and women did not differ in their choice of social comparative referents, including same-gender or downward/upward comparisons, or the choice of their own previous salary as comparative referent. Second, there was no indication that the pattern or strength of relationships between selected social and internal comparative referents and pay allocations differ between the gender groups. These findings suggest that the results of the studies, particularly those using student samples, that did not take into account the effects of paid work experience on women’s and men’s pay allocations must be regarded with caution. Prior to entry into the labor market and in the absence of organizational or occupational pay standards or personal experience with paid work, individuals (e.g., students) may rely on general and impersonal external comparison standards (e.g., beliefs about gender wage gap and its causes in their society). On the other hand, in work settings, employees may rely on both specific external pay standards, which partly mirror gender pay differentials in their organization or occupation, and internal standards, largely their own current or past salaries. To address these speculations (see also Desmarais & Curtis, 1997a, 1997b) and the generalizability of our and previous results, future research may need to apply longitudinal methods to examine effects of changes due to situational factors (e.g., earnings, status, gender and racial composition of occupations) and individual factors (e.g., age, tenure, race, gender, education, and family and economic characteristics and responsibilities). Entitativity and Choice of Comparison Standards Our findings were generally in line with the predictions derived from the entitativity concept (Campbell, 1958) and provide preliminary support for the viability of the cognitive approach to the study of the choice of comparative referents. First, as expected, occupational commonality as an instantiation of common fate in work settings appeared as the best predictor of the choice of comparative referent. Even when occupational commonality was at odds with gender similarity or the motive for upward comparisons, respondents chose a referent within their occupation of dissimilar gender or lower pay. Second, consistent with previous studies, similarity between the comparer and a comparison target predicted the choice of a referent for salary comparisons. Third, in addition to common fate, the relationship between similarity and the choice of comparative referent was also moderated by the availability of comparison targets in five of six participant groups. Finally, the current findings lend weak support for the effect of proximity on the choice of salary referent. Proximity had a statistically detectable effect only when the distance between the comparer and a comparison target was extreme (e.g., one’s own workplace and workplaces abroad). It may be indicative of low effect size of proximity on salary comparisons (cf. Shah, 1998). In our survey, the sources of perceived group entitativity were defined a priori and dichotomously. The results need to be complemented by studies in which the degree of group entitativity is varied experimentally and its effect on the choice and use of comparison standards are measured (cf. Pickett, 2001). In
No. 4:32, 22 summary, the cognitive approach of the perceived group entitativity in combination with other approaches (e.g., social identity and related theories, social network theories) may increase our ability to predict determinants and effects of the selection of comparative referents. More specifically, the perceived group entitativity may further clarify central notions such as salience, attractiveness, and relevance of the referent in previous conceptualizations (e.g., Goodman, 1974; Kulik & Ambrose, 1992; Levine & Moreland, 1987) and facilitate to predict which person or group of persons is perceived as salient or relevant for comparison purposes and under what conditions. With regard to the motives for social comparisons and adjustment of salary allocations, our results showed that respondents demanded higher salary for similar others at the entry level and after five years in their occupation. These results are in accordance with predictions by the self-improvement motive for social comparisons rather than the self-protective motive. The latter predicted that respondents would demand less salary for new colleagues in order to maintain a positive self-esteem by having higher salary as an indicator of success or status. One interpretation of the results is that participants may have perceived a salary increase for new colleagues as an indirect boost to their individual standing through heightened status and salary for their occupation in general. A second complementary interpretation of these results suggests that by demanding higher salaries for less experienced colleagues in the occupation the respondents may have expressed their perception of being underpaid and dissatisfied with their own current salary, which they then can use in negotiation for higher salaries for themselves (cf. Levine & Moreland, 1987). This leads us to a general discussion of the antecedents and effects of comparison standards and their measurements. In line with prior research, our results also indicated that individuals use multiple referents (e.g., Goodman, 1974; Oldham, Kulik, Ambrose, Stepina, & Brand, 1986; for an exception, see Oldham et al. 1982). Furhtermore, the choice of a comparative referent does not necessarily reflect the direct effect and importance of the selected referent on pay allocations and evaluations. For instance, although of all selected referents, participants most frequently selected referents in their occupations within (11.6%) or outside (11.7%) their workplace, there was no direct relationship between the choice of these referents and demanded salaries. By contrast, even though comparisons with salaries of employees within an occupation abroad or their managers constituted lower percentages (5.6% and 4.6%, respectively), the occurrence of such comparisons was directly related to higher salary demands. The weak associations between explicit comparison standards (both social and internal) and salary demands observed in this study are consistent with findings of previous studies that indicated either low to moderate relationships or no relationship between social comparison standards and pay equity or satisfaction at all (e.g., Berkowitz, Fraser, Treasure, & Cochran, 1987; Law & Wong, 1998; Moore, 1991; Oldham, Kulik, Ambrose et al., 1986). Wood (1989) also noted the asymmetry between the choice of social comparative referents and the effect of these comparisons in past research. The limited effects of social comparison standards on outcome variables in this and previous studies may reflect (a) predominantly direct effects and/or (b) measurement limitations. This discussion points to a need for process studies in which indirect (i.e., mediated) and/or
No. 4:32, 23 moderating effects of comparisons standards on pay allocations and related evaluations are examined. Moreover, there is also a methodological aspect to the weak or non-existent relationships between explicit comparison standards and salary demands. For instance, with regard to pay satisfaction, Law and Wong (1998) provided direct evidence for the effect of measurement of social comparisons on observed results, demonstrating that when social comparisons were measured as the frequency of comparisons there was no association with pay satisfaction as indicated elsewhere (e.g., Berkowitz et al., 1987). In contrast, when the measurements of social comparisons and the outcome variable referred to the same dimension (i.e., participants’ satisfaction with own pay compared to a referent’s pay and general pay satisfaction), the association between social comparisons and pay satisfaction was highest. Hence, researchers need pay special attention to the methodological factors in the study of comparison standards and their effects (for a thorough discussion, see Wood, 1996). Managerial Implications The results of this study have direct implications for managements of organizations with large low-paid and/or blue-collar occupations. These organizations, especially in the public sector, face the increasing problem of turnover of personnel and recruiting qualified applicants for job vacancies. In Western European countries, the problem has been accentuated in the public sector since waves of privatization and deregulation have exposed organizations and occupations previously managed by local and national governments to competition from the private sector. With regard to salary, managers of these organizations may choose one of three strategies to handle the turnover and vacancy problems. First, to increase the status and salary of an occupation they may begin to raise starting salaries for new employees and gradually increase salaries for other employees. If we refer to this strategy as a bottom-up process, a top-down process would imply an increase of salary beginning with employees with longest occupational and organizational tenure and working down to new employees. A third strategy, which is financially the least feasible, is the application of both processes simultaneously. In addition to financial consideration, the management needs to predict employees’ reactions to any of these and other compensation-related plans. Our results suggest that employees may support a bottom-up process, and not only tolerate that new co-workers earn more than they did in similar conditions, but initiate such a process. By contrast, if salary decisions are not adequately anchored and supported by experienced employees, the inequity felt may lead to various organizational conflicts, diminished job satisfaction, and increased turnover and absenteeism among employees.
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No. 4:32, 28 Table 2 a Means and Standard Deviations for Variables as a Function of Respondents’ Subsample Membership and Gender Metalworkers’ Union W&R Trade Union Hospital Men Women Men Women Men Women (n = 112) (n = 110) (n = 89) (n = 81) (n = 132) (n = 380) Variable Dependent 14.84 (2.34) 14.74 (2.20) 14.24 (2.29) 12.95 (2.57) 16.04 (2.31) 15.78 (2.32) Demanded starting salaryb Demanded salary after five 18.58 (2.71) 18.03 (2.47) 17.78 (2.71) 16.87 (2.40) 19.71 (3.11) 19.32 (3.18) b years Mediator 7.48 (3.49) Explicit comparison 5.35 (3.26) 4.24 (3.21) 5.06 (3.34) 4.78 (3.57) 7.02 (3.32) standards 17.07 (1.92) 15.21 (2.63) 16.08 (3.15) 13.31 (3.38) 16.71 (2.68) 16.29 (2.81) Comparer’s current salaryb Control 5.79 (1.72) Education 4.08 (1.92) 3.46 (1.90) 4.31 (1.86) 4.28 (1.88) 5.99 (1.50) Job experience 13.46 (10.73) 12.19 (9.07) 14.03 (12.86) 12.30 (9.58) 15.73 (8.91) 19.10 (8.88) aStandard deviations are in parentheses. bSalary measures represent amounts in thousands (Swedish crowns). Table 3 a Comparison Frequency and Percentage as a Function of Comparative Referent’s Occupation, Gender, and Workplace Referent’s occupation Same Other Total b b b f (%) f (%) f (%) % % % Referent’s gender Same 503 (55.6) 8.9 356 (39.4) 6.3 859 (47.5) 15.1 Other 412 (45.6) 7.3 257 (28.4) 4.5 669 (37.0) 11.8 Referent’s workplace Same 657 (72.7) 11.6 427 (47.2) 7.5 1,084 (60.0) 19.1
No. 4:32, 29 Other 663 (73.3) 11.7 436 (48.2) 7.7 1,099 (60.8) 19.4 Abroad 315 (34.8) 5.6 --315 (34.8) 5.6 aPercentage of positive responses (yes) for each cell is in parentheses. bPercentages are computed in relation to a total of 5,670 (100%) observed positive responses to all 13 explicit comparison items.
No. 4:32, 30 Table 4 Comparison Frequency and Percentage as a Function of Respondent’s Gender and Subsample Membership and Referent’s Gender Male Referent Female Referent f (%) f (%) n Binomial or ? 2 Respondent’s Gender Male 175 (52.6) 124 (37.2) 333 40.98*** Metalworkers’ 64 (57.1) 36 (32.1) 112 24.30**a Union W&R Trade Union 49 (55.1) 30 (33.7) 89 (N = 19, k = 0)**a Hospital 62 (47.0) 58 (43.9) 132 (N = 12, k = 4)a Female 288 (50.4) 328 (57.4) 571 17.28*** Metalworkers’ 50 (45.5) 49 (44.5) 110 (N = 11, k = 5)a Union W&R Trade Union 26 (32.1) 38 (46.9) 81 (N = 16, k = 2)*a Hospital 212 (55.8) 241 (63.4) 380 12.85**a aIn six cases, the binomial or McNemar tests are used as post-hoc tests; therefore, probabilities are corrected for six tests. * p < .05; ** p < .01; *** p < .001
No. 4:32, 31 Table 5 a Results of Regression Analyses Predicting Demanded Salary for Others Entering the Occupation and After Five Years in the Occupation Demanded Starting Salary Demanded Salary After five Years 2 b SE b b SE b ∆R ∆R2 Variable β β Step 1: Control .12*** .10*** Education 40.28 .35*** 500.33 49.49 .33*** 441.38 Job experience 19.42 8.00 .08* 7.32 9.83 .02 Step 2: Independent .06*** .03*** Gender -345.46 162.53 -.07* -429.72 203.03 -.07* 134.21 -.28*** W&R Trade cf. all -926.26 167.65 -.20*** 1,046.6 5 Hospital cf. all 745.47 129.88 .25*** 676.34 162.25 .19*** Step 3: Interaction .01** .00 Gender X W&R Trade -815.28 263.98 -.15** -411.69 331.24 -.06 Step 4: Social comparison .03*** .04*** standards Same-occupation, same-gender -477.70 232.57 -.10* -300.90 291.52 -.05 † Other-occupation, same-workplace 340.68 186.29 72.48 233.51 .01 .07 Same-occupation, abroad 463.60 176.76 .09** 700.43 221.56 .11** Highest salary in the occupation 251.17 207.62 .05 570.15 260.24 .09* Lowest salary in the occupation -384.44 191.04 -.07* -606.81 239.47 -.10* .09** 608.36 233.53 .09** Salary of managers 511.20 186.31 Step 5: Internal comparison .11*** .12*** b standards Previous salary -402.02 159.44 -.08* -388.40 199.51 -.06† Current salary 0.31 0.03 .37*** 0.40 0.03 .39*** aN = 904. For each step, only those added variables that contributed significantly to either regression equations are presented.
No. 4:32, 32 bFor
the final model (Step 5), adjusted R2 = .32 for demanded starting salary and adjusted R2 = .28 for demanded salary after five years. † p < .10 * p < .05 ** p < .01 *** p < .001
