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The Impact of Minimum-Wage Increases: Evidence from Fast-food Establishments in Illinois and Indiana

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Abstract

Fast-food establishments in Illinois and Indiana were surveyed during a period of state-mandated minimum-wage increases in Illinois. While entry-level wages of Illinois establishments rose substantially in response to the mandated increases, there is little evidence that Illinois establishments ameliorated wage increases by delaying scheduled raises or reducing fringe benefit offerings. There is little evidence of ‘labor-labor’ substitution in favor of women, better educated, or teenaged workers, or increased worker tenure at the new wage, but weak evidence of increased food prices. In contrast, there are large declines in part-time positions and workers’ hours in Illinois relative to Indiana. Aggregate figures from the Bureau of Labor Statistics support relative declines in total fast-food employment in ‘downstate’ Illinois counties, as hypothesized. However, establishments’ responses do not appear proportionate to the strength of the minimum wage change.

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Notes

  1. The federal minimum wage was $5.15 from 1997 until July 24, 2007, when it increased to $5.85. Like many states, Illinois allows a discount for teenage workers under age 18 of 50 cents from the prevailing hourly minimum and it permits a 30% training discount for up to 2 weeks at fast-food establishments.

  2. Card and Krueger went on to write a book (Card and Krueger 1995) and several more articles about minimum-wage policy. Perhaps most notable is Card and Krueger (2000), which is both a reply to Neumark and Wascher (2000) and an original analysis of confidential, establishment-level BLS data on the firms in their sampled counties in New Jersey and Pennsylvania. This paper tends to focus on Card and Krueger (1994) because it replicates their strategy, but it is certainly not the authors’ last word on this topic.

  3. Specifically, they find increases in job tenure and the share of full-time employees at fast-food restaurants. They also find a price increase at fast food restaurants.

  4. Fractional reports of ‘full-time’ positions are another sign of data irregularities (Welch 1995).

  5. Payroll data were obtained with the help of a restaurant lobbying group. Some have argued that this may introduce response bias.

  6. The median wage of fast-food cooks in the counties of the Chicago Metropolitan area was $7.56 in the first quarter of 2004. For all food-preparation and service workers, including fast-food but also tipped workers, it was $6.65 (figures from the Illinois Department of Employment Security).

  7. CK included Burger King, Wendy’s, Roy Rogers, and Kentucky Fried Chicken.

  8. The questionnaire is available from the author upon request.

  9. Response bias resulting from establishments more adversely affected by the minimum wage being more eager to participate in the survey is unlikely. The survey was administered to the manager on duty, and there was no way for establishments to coordinate responses over time.

  10. The lower share of McDonald’s establishments among respondents is only significant at the 80% confidence level.

  11. I am grateful to a referee for suggesting an analysis of these data.

  12. In 2004 and 2005, the BLS does not disclose sector-specific employment data for 45% of in-sample Illinois counties and 26% of in-sample Indiana counties. In 2003, the BLS does not disclose sector-specific employment data for 25% of in-sample Illinois counties and 42% of in-sample Indiana counties.

  13. FTE-hours is computed as establishment weekly hours of nonsupervisory workers divided by 35, a standard cutoff for full-time work. FTE-positions (the primary measure used by CK) are computed as the sum of reported full-time nonsupervisory positions and one-half of reported part-time nonsupervisory positions. Supervisory positions are added to FTE-positions to compute total FTE-positions. To avoid repetition, throughout the narrative both ‘FTE-positions’ and ‘FTE-hours’ are understood to refer to nonsupervisory workers unless otherwise noted.

  14. The patterns for full-time, entry-level wages (not pictured) are similar.

  15. Of course, as NW and Card and Krueger (2000) note, these comparisons are sensitive with respect to the assumption that 35 hours per week is the full-time worker cutoff. For the chosen cutoffs, FTE-hours appears more variable than NW’s, while FTE-positions appears less variable than CK’s.

  16. Also following NW, when the 2003–2005 change in FTEs is predicted using 2003 FTEs, the coefficients are negative and similar (around −0.35) for both FTE-hours— and nonsupervisory FTE-positions-based estimates, again consistent with the hypothesis that measurement error is no more a problem for positions— than hours-based variables.

  17. NW note that hours could be more variable than employment in the presence of fixed costs to hiring and firing. Overall, the data do not support this hypothesis either.

  18. Closings are rare and generate large employment changes that may unduly influence the findings. Closings are ignored in the analysis.

  19. Log-difference specifications (defined as the ln(Ypost/Ypre)) were also estimated. This specification change makes little qualitative difference to the findings. For both the D-in-D and alternative models, when significant in both the difference and log-difference specifications, findings are somewhat stronger for the log-difference specification (not reported).

  20. Respondents were not asked if the establishment is company-owned. While they were asked if the owner works in the establishment, there are many nonresponses and this information is not used.

  21. Food choices vary greatly and several chains do not offer breakfast. Chains also vary greatly in size. The average number of open registers after 11 a.m. ranges from just 1.2 at Subway to 3.9 at McDonald’s. The number of nonsupervisory workers ranges from just 10.3 at Subway to almost 40 at McDonald’s.

  22. Information on the value of fringe benefits was not collected.

  23. Fewer establishments report hours than positions information. The findings for positions variables are always quite similar when estimated for the subsample of establishments reporting hours.

  24. Findings on workforce composition are not presented in Table 4 but are available upon request.

  25. When the estimation is carried out for the smaller sample of stores that do not report a change in payroll period, estimates are negative and similar in magnitude to those from specifications including chain dummies in Table 5 for 2003–2005 and are negative in sign but not significantly different from zero for 2004–2005.

  26. This case study (two states, each measured ‘before’ and ‘after’) is a classic example of the “two by two” cases that feature prominently in the economics literature. As Donald and Lang (2007) state, “the main feature of this case is that we cannot calculate the standard error of the estimate and thus must exercise considerable caution in drawing conclusions” in the presence of spatially correlated errors. I am grateful to a referee for pointing this out.

  27. Katz and Krueger (1992) suggest using Gap as an instrument for the actual wage change. In this case the percentage change in the actual wage and Gap for Illinois are almost perfectly correlated, since nearly all Illinois establishments were forced to an entry-level wage of exactly $6.50 in 2005.

  28. This is just another way of demonstrating that the statutory gap and the actual wage change are almost the same for all the Illinois firms in the sample.

  29. The estimate for FTE-hours is given in Table 6. The estimate for part-time positions (not given in table) is from a robust regression with chain and payroll period dummies and has a standard error of 1.25.

  30. Price adjustments are path-dependent, infrequent, apparently scheduled (i.e., occurring at pre-determined intervals), staggered across products (i.e., products have their ‘turn’ at a price change), and tend to cluster at certain values. Specifically, MacDonald and Aaronson (2006) find that the impact of the minimum wage appears to be manifested in the removal of products from a ‘sale.’ In these low-frequency data, there is little context for the observed prices and their changes.

  31. Effects on hours per worker are always estimated to be insignificant.

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Acknowledgments

This project would not exist without Ron Baiman and Joe Persky, who first approached me about collecting original data on the Illinois minimum wage change. Their participation in the survey design, surveyor training, and data collection efforts is greatly appreciated. The Russell Sage Foundation provided generous financial support for this project. Students at the University of Illinois at Urbana-Champaign and the University of Illinois at Chicago conducted the surveys of fast-food outlets. Veronica Alaimo and Maghaisvarei Sellakumaran provided excellent graduate research assistance and Kathleen Dorrestein ably assisted with data entry. Alan Krueger and David Neumark provided help and advice at early stages of this work. Nick Powers provided helpful comments on numerous drafts. I am grateful to an anonymous referee for many helpful comments that greatly improved the paper. Responsibility for errors rests with the author alone.

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  1. University of Illinois at Urbana-Champaign, IGPA, 1007 W. Nevada St., Urbana, IL, 61801, USA

    Elizabeth T. Powers

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Powers, E.T. The Impact of Minimum-Wage Increases: Evidence from Fast-food Establishments in Illinois and Indiana. J Labor Res 30, 365–394 (2009). https://doi.org/10.1007/s12122-009-9068-3

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