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Published in final edited form as: J Clin Nurs. 2010 Nov 4;20(1-2):60–67. doi: 10.1111/j.1365-2702.2010.03456.x

The effect of unions on the distribution of wages of hospital-employed registered nurses in the United States

Joanne Spetz 1, Michael Ash 2, Charalampos Konstantinidis 3, Carolina Herrera 4
PMCID: PMC11062236  NIHMSID: NIHMS1987054  PMID: 21054601

Abstract

Aims and objectives.

We estimate the impact of unionisation on the wage structure of hospital-employed registered nurses in the USA. We examine whether unions have an effect on wage differences associated with race, gender, immigration status, education and experience, as well as whether there is less unexplained wage variation among unionised nurses.

Background.

In the past decade, there has been resurgence in union activity in the health care industry in the USA, particularly in hospitals. Numerous studies have found that unions are associated with higher wages. Unions may also affect the structure of wages paid to workers, by compressing the wage structure and reducing unexplained variation in wages.

Design.

Cross-sectional analysis of pooled secondary data from the United States Current Population Survey, 2003–2006.

Method.

Multivariate regression analysis of factors that predict wages, with models derived from labour economics.

Results.

There are no wage differences associated with gender, race or immigration status among unionised nurses, but there are wage penalties for black and immigrant nurses in the non-union sector. For the most part, the pay structures of the union and non-union sectors do not significantly differ. The wage penalty associated with diploma education for non-union nurses disappears among unionised nurses. Unionised nurses receive a lower return to experience, although the difference is not statistically significant. There is no evidence that unexplained variation in wages is lower among unionised nurses.

Conclusions.

While in theory unions may rationalise wage-setting and reduce wage dispersion, we found no evidence to support this hypothesis.

Relevance to clinical practice.

The primary effect of hospital unions is to raise wages. Unionisation does not appear to have other important wage effects among hospital-employed nurses.

Keywords: hospitals, industrial relations, nursing, wages

Introduction

Unions that represent health professionals have become increasingly important in the employment and political landscape of the USA. Nearly 21% of registered nurses (RNs) in the USA were unionised in 2003 (Terlep 2006). In recent years, nursing unions have claimed credit for wage increases across the USA, hospital worker safety improvements, implementation of minimum nurse-to-patient ratios in California, and the introduction of federal minimum staffing legislation in Congress (California Nurses Association 2009). Hundreds of research studies have established that unionised workers in any sector earn more than their non-union counterparts (Lewis 1986), and this finding has been corroborated in studies of nursing employment (Link & Landon 1975, Fottler 1977, Feldman & Scheffler 1982, Bruggink et al. 1985, Hirsch & Schumacher 1998).

In addition to raising wages, unions may also affect the structure of wages paid to workers. Freeman and Medoff (1984) argue that unions compress the wage structure as an expression of solidarity and worker preferences. In this paper, we examine the impact of unionisation on the wage structure of hospital-employed RNs in the USA. We focus on two factors: wage differentials associated with age, gender, education and experience; and unexplained variation in wages, measured as residual wages. We find little evidence that unions have an effect on the wage structure of RNs.

Background

In the past decade, there has been resurgence in union activity in the health care industry in the USA, particularly in hospitals. Of 2·4 million RNs employed in the USA in 2003, unions represented about 472,000, or nearly 21% (Terlep 2006). This rate is over five percentage points higher than for all workers in the USA (Clark & Clark 2006). Membership of hospital-employed RNs in unions in the USA was stable at about 18% from 1983 through the mid-1990s, and grew rapidly since that time (Fig. 1). The apparent volatility in the rate of unionisation reflects sampling variation; the error bands are wide enough to make the spiky series consistent with the smoothed results, which are available from the authors on request. During the first six months of 2000, health care unions, including unions for nurses, service workers and other healthcare employees, won organising elections at a rate of 61·7%, compared with a 33·9% rate for manufacturing (Forman & Davis 2002).

Figure 1.

Figure 1

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Per cent of hospital-employed registered nurses who are union members.

Unions represent the collective interests of employees in negotiating the terms and conditions of employment with employers and thereby affect many aspects of the employment relationship. Unions typically seek better working conditions for their members, including higher wages, more employment security and improved terms of hiring, promotion and layoff (Baumol & Blinder 1991). They bargain with employers over the distribution of profits (or net revenues in the case of non-profit or public entities). Many studies of workers in all industries have found that unions are associated with higher wages and fringe benefits (Mellow 1979, Lewis 1986, Freeman & Kleiner 1990, Jakubson 1991, Wunnava & Ewing 1999, Hirsch & Schumacher 2001). Research on nursing employment also has found that unionisation is associated with higher wages (Link & Landon 1975, Fottler 1977, Becker 1979, Sloan & Steinwald 1980, Feldman & Scheffler 1982, Bruggink et al. 1985, Hirsch & Schumacher 1995, 1998). Some studies find that unions affect wages received by non-union workers (Lewis 1990, Hirsch & Schumacher 1998).

Because unions bargain for wage contracts that affect all nurses in their bargaining unit, they also may address differences in earnings that arise from differences in seniority, education or other worker characteristics (Freeman & Medoff 1984). Unions may increase some earnings differences, such as by negotiating career ladders that provide automatic returns to seniority, and collapse others, such as by limiting the importance of supervisor evaluations in determining pay. Freeman (1980) reported that unions contributed to reductions in wage differences between white-collar and blue-collar workers in firms. In a later study, Freeman (1982) again found that within-employer wage dispersion was much smaller in unionised establishments, and attributed this to intentional differences in wage practices. For example, unionised employers were more likely to increase wages based on automatically progressing scales rather than based on individual merit reviews. Other studies (Hirsch 1982, Belman & Heywood 1990, Lemieux 1998) also found that unions reduced wage differences associated with observed measures of employees’ skills.

Unions may have an ambiguous effect on the wage structure of nursing. On the one hand, standard solidaristic pay-compression arguments may dominate. On the other hand, if hospitals exert significant monopsony power in nursing, then union wage strategy may address the monopsonistic pay structure (Yett 1975, Sullivan 1989, Staiger et al. 2010). Under monopsony, employers focus as much as possible on attracting the ‘marginal nurse’, e.g. with hiring bonuses or other disproportionate escalation of starting salaries. Because monopsonistic firms focus on the marginal worker while unions represent the interests of all workers, union strategy in the face of a monopsonistic pay structure may attempt to reward nurses with more job attachment. Thus, nurse advocates prominently criticise the lack of career ladders or skill-based pay scales (Queneau 2006), the absence of which is consistent with monopsonistic strategy on the part of the employer. In addition to the monopsonistic explanation, other advocates have attributed limited career ladders to traditional feminisation of the profession (Apesoa-Varano & Varano 2004). Both the gender and monopsonistic explanations are consistent with weak rewards to experience resulting in high turnover and exit. The introduction or steepening of career ladders tends to decompress the wage structure for nurses. The predicted union effect on the wage structure is thus ambiguous.

Methods

Data

Our source of data on nurses is the merged outgoing rotation groups of the 2003 through 2006 Current Population Surveys (CPS), which is collected by the United States Bureau of Labor Statistics. Union membership data were first collected on a regular monthly basis beginning with the 1983 CPS. From the CPS, we extracted all RNs (Census Occupation Code = 3130) working in hospitals (Census Industrial Code = 8190) whose incomes were directly reported (and not imputed). Although the CPS is in principle a longitudinal dataset, with each participant appearing twice in the outgoing rotation group, we ignore the repeated observations, both in terms of identification (Hirsch & Macpherson 2000) and in terms of clustered errors. The CPS sample used for this analysis, after applying restrictions described later, had 831 RNs who were unionised and 3403 non-union nurses. Because we pool nurses from multiple years of the CPS, there is not a straightforward comparison of the sample size to the national population of RNs, but the unionisation rate and the share of RNs employed in hospitals is consistent with other data sources, such as the 2004 National Sample Survey of Registered Nurses (Health Resources and Services Administration 2006).

The hourly wage for RNs is computed as usual weekly earnings divided by usual weekly hours. We use the CPI-U-RS, a time-consistent measure of the US consumer price level, to convert all values to real 2006 dollars. Educational attainment was based on credentials received by nurses; however, we cannot observe whether the completed degree was in the field of nursing. We code nurses with no degree beyond secondary as Diploma nurses and other nurses by the highest degree attained. We used potential experience – i.e. age minus six minus years of education – as a proxy for years of work experience.

We applied the following additional restrictions to our sample. In addition to requiring that all necessary variables be available for all observations, we restricted the sample to nurses who usually worked at least 20 hours per week, and whose real hourly earnings fell between the Federal minimum wage and $100 per hour (in 2006 dollars).

Analysis

To measure the impact of unions on the wage structure of nurses, we estimate two multivariate ordinary least squares equations – one for unionised nurses and one for non-union nurses. The dependent variable in these equations is the natural logarithm of computed hourly wage. The explanatory variables include human capital and demographic characteristics: education, experience and experience squared, gender, race/ethnicity, and immigrant status. We also control for the region of the USA where the nurse lives, which is measured with a set of dummy variables. There are nine regions, each of which we subdivide into two sub-regions – one for rural residents and one for urban residents. Thus, we have 18 regions, represented by 17 dummy variables in the equations (urban northeast is the excluded region). Finally, we include three dummy variables to control for each year of data, with 2003 as the omitted year.

After estimating the two equations, we first compare the values of key coefficients between these equations using t-tests to learn whether there are significant differences in the effect of selected variables on wages. We focus on how unionisation status affects pay differences that are widely observed: gender, racial/ethnic, educational and seniority. Second, we examine wage variation and residual wage variation by union status. We compare the R-squared values of the equations, which measure the amount of variation in the data explained by the wage equations. We also compare the distributions of residual wages from the equations, thus explicitly assessing whether there is greater unexplained variation between union and non-union RN wages.

Results

Table 1 presents characteristics of union and non-union RNs employed in hospitals, with data from 2003–2006 combined. Average hourly earnings are substantially higher for unionised nurses. Union and non-union RNs are both overwhelmingly female, and their age and educational profiles are similar, with slightly more Associate Degree nurses in the non-union group. Unionised RNs are much more likely to be immigrants (18·9 vs. 10·1%) and more likely to be non-white (28·1 vs. 17·5%). They also are more likely to live in urban locations (85·6 vs. 80·0%). These data suggest that wage differentials between union and non-union RNs could arise from differences in location, race and education, as well as from the impact of unions themselves.

Table 1.

Characteristics of union and non-union registered nurses employed in hospitals, 2003–2006

Union Non-union
Real hourly earnings $33·50 $28·20
Female 91·1% 92·4%
Immigrant 18·9% 10·1%
Urban location of residence 85·6% 80·0%
Non-hispanic white 71·9% 82·5%
Black 6·7% 6·0%
Hispanic 5·2% 3·5%
Other 16·2% 8·0%
Diploma 4·7% 4·7%
Associate’s degree 37·1% 40·4%
Bachelor 50·4% 49·0%
Master’s 5·4% 4·3%
Doctorate 2·3% 1·6%
Age 42·85 41·17
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Table 2 reports coefficients from the ordinary least squares regressions of the natural logarithm of real hourly wage on human capital and demographic characteristics for unionised and non-unionised RNs. Among unionised nurses, wages are higher for nurses with bachelor’s and master’s degrees, when compared to those with associate degrees. As expected, wages increase with experience, but at a diminishing rate. There are no differences in wages associated with gender or immigrant status, and the only race/ethnicity gap is for Hispanic nurses, who earn a lower wage than non-Hispanic white nurses.

Table 2.

Coefficients from multivariate regression equations for registered nurse (RN) wages, 2003–2006 (standard errors in parentheses)

Unionised RNs Non-union RNs Difference (union–non-union)
Female −0·030 (0·037) −0·005 (0·024) −0·026 (0·044)
Immigrant 0·020 (0·048) −0·064** (0·030) 0·084 (0·057)
Race/ethnicity (non-hispanic white omitted)
 Black −0·035 (0·069) −0·068** (0·032) 0·033 (0·076)
 Hispanic −0·108* (0·058) −0·042 (0·042) −0·066 (0·071)
 Other 0·024 (0·046) 0·043 (0·034) −0·019 (0·058)
Education (Associate’s degree omitted)
 Diploma 0·075 (0·060) −0·053* (0·028) 0·129** (0·066)
 Bachelor 0·083** (0·027) 0·074** (0·012) 0·008 (0·029)
 Master’s 0·237** (0·053) 0·197** (0·033) 0·040 (0·063)
 Doctorate 0·063 (0·087) 0·045 (0·047) 0·018 (0·099)
Potential experience 0·017** (0·005) 0·019** (0·002) −0·002 (0·005)
Potential experience squared −0·0003** (0·0001) −0·0003** (0·00004) −0·00002 (0·0001)
R-squared 0·27 0·19
n 832 3404
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Dummy variables are included to control for year and for region.

*

Significant at 0·10.

**

Significant at 0·05.

The second column of Table 2 provides coefficients for non-union nurses. As with unionised nurses, wages are higher for nurses with bachelor’s and master’s degrees; however, among non-union nurses, diploma-educated nurses experience a wage penalty when compared to those with associate degrees. Wages increase with experience, but at a diminishing rate. There is no gender difference in wages. Non-union immigrant nurses and black nurses earn significantly less than do non-immigrant and non-Hispanic white nurses.

The last column of Table 2 provides the differences between the coefficients of the union and non-union equations, and the standard errors of these differences. The standard errors are generally large. The only coefficient that is significantly different between the union and non-union equations is that for the diploma-education indicator. Unionised diploma nurses earn substantially more than do non-union diploma nurses. Note that the R-squared values of the equations are quite different; 27% of variation in wages is explained in the union equation, while only 19% of variation is explained in the non-union equation.

The age-earnings profile in the union sector is presented in Fig. 2. Because the age-earnings specification is quadratic in potential experience, the most straightforward way to interpret the difference is through the plot of the predicted experience premium. The profile is somewhat flatter than in the non-union sector, which suggests that unions might compress earnings of older and younger nurses. The difference in profiles accounts for approximately 5% of wages at their peak at approximately 30 years of experience. The flatter wage profile in the union sector runs contrary to the hypothesis that career ladders are a union goal. The result is, however, consistent with union compression of pay scales. The result is also consistent with differential attrition in the union and non-union sectors: if high pay is required for the long-term retention of nurses in the non-union sector, then the steeper age-earnings profile for non-union nurses might simply indicate that all of the lower-wage nurses exit the field over time.

Figure 2.

Figure 2

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Experience-earnings profiles of union and non-union registered nurses, 2003–2006.

We examine the overall distribution of wages in the union and non-union sectors. The standard deviation of the logarithm of wage is a standard measure of dispersion. The non-union and union sectors have very similar standard deviations of log wage, 0·305 and 0·326, respectively, which suggests that wage compression is not a strong force in the union sector. The similarity persists in the standard deviation of residual log wage or the dispersion of wages after accounting with regression for characteristics that typically determine wage differences. The standard deviation of residual log wage – or the standard error of the regression – is 0·275 in the non-union and 0·280 in the union sector. Histograms of residual log wage in the two sectors appear in Figs 3 and 4; the similarity in the distributions suggests that wage compression is not an important feature of unions in nursing.

Figure 3.

Figure 3

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Histogram of residual log wages for non-union registered nurses, 2003–2006.

Figure 4.

Figure 4

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Histogram of residual log wages for union registered nurses, 2003–2006.

Discussion

This analysis provides mixed evidence regarding whether unions have an effect on the wage structure of RNs. We find little evidence that there are differences in wages associated with gender, immigrant status, race/ethnicity, education or experience. Although wage equations indicate that immigrant and black non-union RNs face a wage penalty, while unionised immigrant and black nurses do not, the differences in these coefficients between the union and non-union equation are not statistically significant. We also find little evidence supporting theories about the role of unions in wage rationalisation.

The nursing workforce is relatively homogeneous when compared with other professions; all nurses have reached an educational standard and hold a licence to practice, and it is a predominantly female profession (Queneau 2006). Nickell (1977) notes that in the long term, wage differentials among a homogeneous group of workers should be related solely to differences in employer characteristics such as working conditions, and products or services provided. In the case of nurses, it is arguable that there should be little wage dispersion, and thus it is unlikely that unions could have an impact on the wage structure. In an analysis of blue-collar workers, Macpherson and Stewart (1987) found little evidence that unions have an impact on the dispersion of wages of women. In his study of Australian workers, Borland (1996) associated declines in union density with increased earnings dispersion among males, but not among female workers. Our work found no significant associations between unionisation and the impact of education and gender on wages, with the exception of a wage penalty for non-unionised diploma nurses. This penalty may be linked to occupational skill level.

Occupation or skill level also might be associated with the impact of unions on the wage structure. Using disaggregated data from the United States Current Population Survey, Card (1996) determined that unions increase wages more for workers with lower skill than for those with greater skill. Although our study did not directly measure nursing skill, it did examine nursing experience and nursing education, with results that have some consistency with those of Card (1996). We found that non-unionised diploma nurses suffered a significant wage penalty, whereas unionised diploma nurses did not.

Unions may be protective for workers who otherwise might face a wage penalty. Several studies have demonstrated that discrimination is more likely to occur in less competitive or monopolistic industries. Peoples (1994) examined racial wage differentials and found that racial wage gaps were larger among non-union workers in non-competitive industries, when compared with competitive industries. In a later study, Agesa and Monaco (2006) demonstrated that the wage gap between whites and minorities was larger in monopolistic industries than in competitive industries for non-union workers and that unexplained wage residuals were smaller in competitive industries. Their analysis suggested that unions protect workers from the discrimination observed in monopolistic industries. To the extent that the hospital industry is non-competitive or monopolistic, as some research has suggested (Abraham et al. 2005), we would expect the racial wage gap to be smaller among unionised nurses. Indeed, in our analysis, there are wage penalties for non-union black and immigrant nurses, while racial wage differences are insignificant at the 5% level among unionised nurses.

Unions may rationalise wage-setting, leading to fewer unexplained differences in pay. This hypothesis is not supported by an examination of the residual log wages from the regression equations. However, there is a modest indication of more transparently structured pay in the union sector: the R-squared for the non-union sector is 19%, while the R-squared for the union sector is 27%. These results are only weakly consistent with those of the classic work of Freeman (1980 and 1982), who found strong evidence that unions reduce wage dispersion and rationalise the wage structure.

Our focus on a single workforce and economic sector also may explain the different findings between this study and previous work. In previous studies of the impact of unions on wage dispersion, researchers have typically examined cross-industry data or cross-occupation data. For example, Freeman (1980) focused on differences between blue-collar and white-collar workers in firms. Subsequent research has demonstrated that there may be differences in the effect of unions on the wage structure according to industry and occupation (Freeman 1982, Hirsch 1982, Macpherson & Stewart 1987, Peoples 1994). More work needs to be carried out on single-sector unionisation and on the differences between manager and non-manger nurses in a unionised environment.

Conclusion

Wage differentials can reflect rewards provided to nurses with greater skill and knowledge; however, they also might be associated with discrimination and monopsonistic wage-setting practices. In addition to striving to raise wages, hospital unions also might seek to change the variation in wages. We find weak evidence that unions reduce wage differences associated with race and immigration status and some additional evidence that wages are more closely associated with the collective set of observable factors among unionised nurses. We also find some indication that the wages of unionised nurses rise more slowly with experience than do the wages of non-union nurses.

Relevance to clinical practice

Unions are a method of exercising collective voice about the workplace. Some RNs may want steeper career ladders to reward skill and to combat traditional stereotypes of nursing as a profession without opportunities for advancement. Other RNs may prefer pay compression to support professional solidarity. The collective representation provided by unions can reflect these complex preferences. The mixed empirical results are consistent with this complexity and certainly call for further and more detailed study of the micro-operation of nurses’ unions.

Nurse interest in unions may grow if nurses believe their wages are set unfairly. To the extent that the race and immigrant wage differences we measured among non-union nurses are real and perceived by nurses, there is motivation to seek remedy to such discrepancies. This is one of many factors driving increased unionisation among nurses (Clark et al. 2001). Unionisation efforts have been particularly successful in hospitals and among nurses. Union elections are more likely to be successful in workplaces where women account for 75% or more of the workforce, women are in leadership positions, and women of colour are in positions of influence (Bronfenbrenner 2005). The relatively homogeneous, cohesive nature of the nursing workforce naturally lends itself to strive to ensure that there are equitable wages in the workplace.

Acknowledgements

The authors acknowledgement support from Grant-in-aid for Scientific Research (B: principal investigator Yoshifumi Nakata) by The Japan Society of Promotion of Science and Grant for International Collaborative Research (principal investigator Yoshifumi Nakata) by Pfizer Health Research Foundation. The authors thank Jean Ann Seago and Jennifer Kaiser for their support and comments in the process of writing this paper.

Footnotes

Conflict of interest

The authors have no competing interests associated with this manuscript.

Contributor Information

Joanne Spetz, School of Nursing, University of California, San Francisco, CA.

Michael Ash, Economics and Public Policy, University of Massachusetts, Amherst, MA.

Charalampos Konstantinidis, Department of Economics, University of Massachusetts, Amherst, MA.

Carolina Herrera, School of Nursing, University of San Francisco, CA, USA.

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