Abstract
Large, low-cost entrants can have a disrupting impact on incumbents. Supercenters may influence competing supermarkets’ prices, but their effects on competition between supermarkets’ private labels and manufacturer brands remain poorly understood. This study examines the impact of Wal-Mart Supercenters on supermarkets’ strategic use of private labels and the control they may exercise over the pricing of manufacturer brands. We use a structural model to investigate alternative pricing scenarios in which the supermarkets have varying control over manufacturer-brand retail prices. The analysis is applied to the milk market in the Dallas/Fort Worth metroplex in an early period of Supercenter growth: 1996–2001. We find that Wal-Mart induces supermarkets to price their private label more competitively and has a very small effect on the pricing of the manufacturer brand. We also identify the type of consumers Wal-Mart Supercenters can attract from supermarkets, leaving incumbents to face a different composition of consumers. Consumers that continue shopping at traditional supermarkets are found to be less price-sensitive, view incumbents as less heterogeneous, and experience the manufacturer brand as more differentiated from private labels.
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Notes
Wal-Mart was re-branded as Walmart and the Supercenter name was phased out in 2008; we use the original branding because: (1) it is contemporaneous with the time period of analysis and (2) we focus only on Wal-Mart stores selling a full line of grocery products.
The effects of WMS on the conduct of upstream firms have not been investigated mainly due to a lack of data, which, with current methodologies, renders the investigation of strategic interactions via a structural model of the vertical channel infeasible.
Our model is one of horizontal competition between brands, and we do not directly study supply chain effects.
Using data from 1994 to 2006 Ellickson and Grieco (2013) show that WMS’ entry had a lesser impact on the food retailing sector than Wal-Mart’s Discount Stores had on the general merchandise sector.
Our case study focuses on milk, and our findings may not apply to other products.
An early hypermart was converted in 1990.
Hypermart USA primarily failed due to the costs associated with its huge footprint, which exceeded 220,000 sq ft (vs the approximately 178,000 sq ft of a Supercenter) and functioned as mini-malls and included food courts.
Even thirty years later, understanding the effects during this initial period of WMS’ growth is difficult. In recent years, Wal-Mart has disclosed location and sales data to market research firms. However, information on these early years has still not been released.
Anecdotally, a Dallas/Fort Worth newspaper reported that although Kroger claimed to be the first to cut prices which “corresponded to a slight decline in raw milk prices,” it was more motivated by providing value to their customers and increasing volume at its local dairy (Press 1999). The same article describes Minyard and Albertsons responding to Kroger’s price cut by lowering their own prices (Press 1999).
The correlation between one-, two-, and three-month lags is, respectively, 0.1216, 0.1289, and 0.1270 indicating that retail price correlation with lagged farm milk price is also low. These are smaller than values observed by researchers studying other markets (e.g., Yu and Gould 2019; Carman and Sexton 2005).
The result is similar if we transform the variables to logs, with a statistically significant raw milk price coefficient of 0.43 and an \(R^{2}\) of 0.0198.
Given the lack of data on WMS, we treat WMS as one of the purchase decisions in the outside option.
The ways in which we modify the model to account for the expansion of WMS are ad hoc.
For example, in the absence of WMS, incumbent stores may be crowded and, as consumers leave incumbents to shop at WMS, incumbent stores can become less crowded (which can mean shorter lines and more attention from store employees, at least in the short run).
We are allowing product and consumer characteristics to interact only among factors common to each brand and retailer, respectively, similar to using fixed effects.
We tested whether \(\sigma _{J}\) and \(\sigma _{I}\) satisfy the theoretical bounds (to be between 0 and 1) within the range of the data and found that they did.
The ownership matrices corresponding to the pricing scenarios are fully illustrated in the Appendix in Tables 9–11.
These data are unique in that they provide quantity and prices for complete market coverage of grocery retailers, excluding only WMS (i.e., all retailers opted in to inclusion which is not true of all IRI retailer data sets available). Their use in research outputs also does not require approval by IRI, as use of more nationally representative sets that are available do.
Fringe manufacturer brands and leading national brands may behave differently and aggregating these two together may have implications for our results, namely, if the leading national brand has a large share of the market, it may underestimate the degree of power of the leading national brand by diluting its effect with the fringe. This is an empirical issue and in the specific case studied here Borden has a share of approximately 14% of aggregated incumbent volume sales; the next largest manufacturer brand is Morningstar with about 9% of incumbent volume sales; the remaining brands comprise about 5% of incumbent volume sales. Further, each manufacturer brand’s share of volume sales in each retailer (excluding Minyard) is less than 10%, including Borden.
Each retailer’s store brand is sold exclusively via their supermarkets and therefore is not a purchase option at other retailers’ stores.
This was done using the serving size suggested by the USDA Food Guide Pyramid (USDA (1992 - 2005)), which are assumed to be at least proportional to the “true” serving size. The total market size is defined as \(\tau *\mathrm {population}*0.075\) gallons\(*28\) days, with \(\tau \) assumed equal to 1, following Nevo (2001); results are robust to scaling \(\tau \).
Thanks to an anonymous reviewer for highlighting the importance of promotions. See Cleary and Lopez (2014) for use of strategic promotions in this market.
We also tried other modifications of this extrapolation; results are largely robust and are available in the Appendix.
The national number of WMS and W have an unconditional correlation coefficient of 0.95, which is significant at the 1% level. Their correlation remains unchanged during the period of low pricing. The period of low pricing explains less than 10% of the variation in W, and the national number of WMS explains about 90%; together, they explain about 90% as well.
Calculated as (\(\sigma _{0J}+\sigma _{1J}*12)\).
Note that the bounds of \(\sigma \) include 0, so that this finding is consistent with the theory presented in previous sections.
Anecdotally, according to a newspaper report at the time, a spokesperson for Minyard claimed that while they tried to match the prices of other stores, most times it was not feasible. A spokesperson for Tom Thumb said it is part of their business model to “keep milk prices competitive” (Press 1999).
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Rebecca Cleary and Jean-Paul Chavas declare that they have no conflicts of interest. This study was supported in part by the Colorado Agricultural Experiment Station.
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We would like to thank the Zwick Center for Food and Resource Policy at the University of Connecticut for sharing the Dallas/Fort Worth milk market data. We are also grateful to two anonymous reviewers, Alessandro Bonanno, Tasneem Chipty, Vardges Hovhannisyan, Rigoberto Lopez, Kyle Stiegert, Guanming Shi, and members of the Food Research Working Group at the University of Wisconsin for conversations that led to a substantially improved manuscript. We also thank seminar participants at the University of Minnesota, the Department of Justice, the Economic Research Service, and Analysis Group for helpful suggestions. All remaining errors are our own. The data that support the findings of this study are available from the Zwick Center for Food and Resource Policy at the University of Connecticut. Restrictions apply to the availability of these data, which were used with permission for this study.
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Cleary, R., Chavas, JP. Strategic supermarket pricing of private labels and manufacturer brands. Empir Econ 62, 2921–2950 (2022). https://doi.org/10.1007/s00181-021-02123-2
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DOI: https://doi.org/10.1007/s00181-021-02123-2