WORKING 

PAPERS 

Do Retail Mergers Affect Competition? 

Evidence from Grocery Retailing

Daniel Hosken 

Luke M. Olson 

Loren K. Smith 

WORKING PAPER NO. 313 

December 2012 

FTC Bureau of Economics working papers are preliminary materials circulated to stimulate discussion 

and critical comment. The analyses and conclusions set forth are those of the authors and do not 

necessarily reflect the views of other members of the Bureau of Economics, other Commission staff, or 

the Commission itself. Upon request, single copies of the paper will be provided. References in 

publications to FTC Bureau of Economics working papers by FTC economists (other than 

acknowledgment by a writer that he has access to such unpublished materials) should be cleared with the 

author to protect the tentative character of these papers. 

BUREAU OF ECONOMICS 

FEDERAL TRADE COMMISSION 

WASHINGTON, DC 20580
*We would like to thank Pauline Ippolito, Paul Pautler, David Schmidt, Mike Vita, and Matt 

Weinberg and seminar participants at Drexel University for their comments. The views 

expressed in this paper are those of the authors and do not represent those of the U.S. Federal 

Trade Commission or any individual Commissioner. 

Do Retail Mergers Affect Competition? 

Evidence from Grocery Retailing 

Daniel Hosken 

Luke M. Olson 

Loren K. Smith*

U.S. Federal Trade Commission 

600 Pennsylvania NW 

Washington, DC 20580 

December 2012 

Abstract 

This study estimates the price effects of horizontal mergers in the U.S. grocery retailing industry. 

We examine fourteen regions affected by mergers including both highly concentrated and 

relatively unconcentrated markets. We identify price effects by comparing markets affected by 

mergers to unaffected markets using both difference-in-difference estimation and the synthetic 

control method. Our results are robust to the choice of control group and estimation technique. 

We find that mergers in highly concentrated markets are most frequently associated with price 

increases, while mergers in less concentrated markets are most often associated with price 

decreases. 

 
1 

I. Introduction 

Economists have long believed that, other things equal, increases in market concentration 

reduce competition. In turn, less competitive markets lead to higher consumer prices and reduce 

consumer welfare. This belief provides the basis for much of the world’s antitrust policy. The 

U.S., U.K. and E.U., for example, review mergers prospectively. While each agency operates in 

a different legal environment, the economic logic underlying merger review is the same. 

Horizontal mergers can create or enhance market power by combining firms producing substitute 

products.1

 The problem for regulators is determining which mergers are likely to result in 

reduced competition. Unfortunately, there is remarkably little reliable systematic evidence 

linking measures of market concentration such as the Herfindahl-Hirschman Index (HHI) to 

manufacturer markups or consumer prices.2

Empirically identifying a causal relationship between price and market concentration is 

extremely difficult because market concentration is rarely exogenously determined. Demsetz 

(1973) noted firms that attain large market shares are likely those that are most efficient, and that 

markets where scale economies are important will tend to be dominated by a small number of 

efficient firms. As a result, studies that simply estimate the price/concentration relationship 

without controlling for the endogeneity of market structure are unlikely to be successful (Evans 

et al. (1993)), Bresnahan and Reiss (1989, 1990)). 

In this paper we estimate the relationship between consumer prices and market structure 

by examining how prices change following significant changes in market structure resulting from 

horizontal mergers in the supermarket industry. Like other retail industries, the supermarket 

 

1

 See Section 1 of the 2010 U.S. Department of Justice(DOJ)/Federal Trade Commission (FTC) Horizontal Merger 

Guidelines for a clear description of the economic logic underlying U.S. horizontal merger policy. 2

 The Herfindahl-Hirschman Index is defined as the sum of the squared market shares of market participants, where 

firm’s market shares are typically measured as percentage points. 
2 

industry is often viewed as one in which entry and expansion should be relatively easy. 

Assuming a retailer has an existing supply network, it need only identify an effective location 

and obtain permission from local regulators to open an establishment. The sunk costs of entry 

(e.g., the cost of structures and permits), in particular, are likely much lower than in most 

industries. Overall, the industry appears to be quite competitive with estimated net post-tax 

profit margins of only 1.37% over the last decade.3

 Notwithstanding the perceived ease of entry 

and expansion in the industry and relatively low profit margins, mergers in retail markets are 

often subject to material antitrust review. Between 1998 and 2007, for example, the FTC 

investigated supermarket mergers affecting 153 antitrust markets and challenged mergers in 134 

of those markets.4

 Surprisingly, given this enforcement focus, we are aware of only one study 

that has estimated the price effects of a consummated horizontal merger in this industry.5 

We infer causality using two related empirical techniques. We begin by conducting a 

standard difference-in-difference analysis: we compare prices in markets experiencing a merger 

to those in similar markets not experiencing a major change in market structure resulting from 

entry, exit, or a horizontal merger. The major criticism of this method is that the decision to 

merge may be related to market participants’ expectations about future prices in an industry 

resulting in biased price effects (Nevo and Whinston (2010)). To address this concern we 

examine the robustness of our findings to the choice of comparison group and also estimate price 

effects using the synthetic control group approach suggested by Abadie et al. (2010). 

 

3

 Food Marketing Institute estimates of grocery store chain’s net profits, available at: http://www.fmi.org/docs/factsfigures/net-profit-percent-sales-2011.pdf?sfvrsn=2, last visited November 6th, 2012. 4

 Horizontal Merger Investigation Data, Fiscal Years 1996-2007, Federal Trade Commission, Table 4-2. Available 

at http://www.ftc.gov/os/2008/12/081201hsrmergerdata.pdf. 

5

 Huang and Stiegert (2009) estimate the price effects of the merger of grocery retailers Kohls and Copps in 

Madison, Wisconsin. They find that while the merger did not result in a price increase in the months immediately 

following the merger, the merged firms’ prices had risen relative to the control market two years later. 
3 

Overall, our results are supportive of the hypothesis that increases in market 

concentration resulting from mergers cause prices to increase when mergers take place in already 

concentrated markets. In analyzing horizontal mergers, antitrust agencies look at the level and 

change in market concentration associated with a merger as a predictor of competitive harm. 

The 2010 Horizontal Merger Guidelines, for example, state that “Mergers in highly concentrated 

markets [markets with an HHI greater than 2500] that involve an increase in the HHI of more 

than 200 points will be presumed to be likely to enhance market power.”6

 In contrast, mergers in 

unconcentrated markets (with an HHI of less than 1500) resulting in a small change in market 

concentration are viewed as unlikely to be anticompetitive. In this study, we estimate the price 

effects of eight mergers in highly concentrated markets and six mergers in moderately 

concentrated or unconcentrated markets. Our results tend to confirm the presumptions of 

antitrust regulators as stated in the Horizontal Merger Guidelines. We find that five mergers 

resulted in estimated price increases of more than 2% and that four of those were in highly 

concentrated markets. Five mergers resulted in estimated price decreases of more than 2% and 

only one of those occurred in a highly concentrated market, while the remaining four mergers 

were associated with relatively little change in price. These findings are robust to the choice of 

comparison group and estimation technique. 

Our paper contributes to a small but growing literature which estimates the change in 

price following mergers of competing firms. The goal of most papers in this literature is to 

measure the efficacy of antitrust enforcement. In the typical study, researchers identify mergers 

that were likely on the antitrust margin; that is, those mergers that the antitrust authority 

seriously considered challenging but allowed to go forward. If the merger resulted in a price 

 

6

 The 2010 FTC/DOJ Horizontal Merger Guidelines define highly concentrated markets as those having a 

Herfindahl-Hirschman Index (HHI) greater than 2500. 2010 Horizontal Merger Guidelines, 5.3. 
4 

increase, the researchers conclude that policy was too lax while a price decrease indicates the 

regulator may be too strict (Ashenfelter et al. (2009)). The majority of these studies find that 

horizontal mergers increase prices. 7,8 The ability to draw general conclusions regarding the 

efficacy of horizontal merger policy from the published literature, however, is limited. Only a 

tiny fraction of the thousands of mergers filed with the U.S. antitrust agencies have been studied, 

and the majority of those have examined mergers in a handful of industries with a history of 

regulation, e.g., banking, airlines, hospitals, and petroleum. Further, most studies in this literature 

are case studies. While the case study methodology is often essential to credibly identify the 

price effects of mergers (learning and controlling for factors that affect industry pricing), the case 

study approach makes generalization of the finding of any one study to other situations difficult. 

A strength of our study, in contrast to much of the literature,9

 is that we estimate the price 

effects of many mergers affecting different geographic markets with different levels of market 

concentration at roughly the same time. In particular, we estimate the price effects of mergers in 

already highly concentrated markets and examine large mergers in less concentrated markets 

unlikely to result in a reduction in competition. By also studying mergers in less concentrated 

markets, we are able to examine mergers that are likely to be competitively benign but that could 

result in efficiencies which lower consumer prices. Our approach follows Carlton’s (2009) 

suggestion that researchers should examine the price effects of all mergers (those likely and 

unlikely to result in price effects) to more fully understand how mergers affect the competitive 

process. 

 

7

 See Weinberg (2008), Pautler (2003), and Hunter et al. (2008) for recent surveys of this literature. 8

 The major exception is the petroleum industry where the evidence is quite mixed, see Silvia and Taylor (2009) and 

cites therein. 

9

 Prager and Hannan’s (1998) study of banking mergers , Kim and Singal’s (2003) study of airline mergers, and 

Dafny’s (2009) study of hospital mergers for are notable exceptions. 
5 

The remainder of this paper is organized as follows. Section II describes our data 

sources, and Section III presents the methodology used to construct our merger and comparison 

markets. Section IV describes our estimation strategy and presents the empirical findings of the 

study. Section V concludes. 

II. Data 

Our study uses three data sources. The first is A.C. Nielsen’s Trade Dimensions retail 

database. Each year Trade Dimensions creates a census of retail outlets operating in the U.S. for 

a number of retailing industries, including supermarkets, club stores, liquor stores, convenience 

stores, and restaurants. In this study we focus on the primary formats used for grocery retailing: 

conventional supermarkets, supercenters, and club stores.10 Our dataset consists of annual 

observations, including the location, size, estimated sales, the store’s banner (the name the store 

operates under), and corporate ownership of each supermarket, supercenter, and club store in the 

U.S. from 2004 through the fall of 2009. An additional feature of the dataset is that every store 

location has a unique identification number that allows us to track stores over time. For 

example, we can observe if a location changes ownership or if a supermarket that closes for a 

time reopens as a supermarket. The dataset also contains information on the ownership of 

different chains, which is important because many firms operate multiple retail brands, 

sometimes even within a relatively small geographic area. As we describe in the next section, 

this data allows us to identify the entry and exit of retailers from local markets and identify the 

merger of retailers. 

 

10 We exclude other retail formats in the Trade Dimensions Grocery dataset – limited assortment, natural/gourmet 

food, warehouse, and military commissary – because they are so differentiated from traditional supermarkets. For 

example, of these retail formats, only military commissaries offer one-stop-shopping. However, military 

commissaries are available to only a subset of the population. 
6 

The price data we use consists of the prices used to construct the ACCRA Cost of Living 

Index, which is published by the Council for Community and Economic Research (CCER). The 

ACCRA price index is designed to compare the cost of living for moderately affluent 

professional and managerial households in different U.S. metropolitan areas at a point in time.11 

The price data assembled by CCER are collected by the staffs of the roughly 350 local U.S. 

Chambers of Commerce who participate in the data collection project. In the first, second, and 

third quarter of each year, staff of participating Chambers of Commerce collect price quotes for 

60 distinct products corresponding to broad categories of consumer expenditures, including 

housing, energy, food, transportation, and health care.12 In this study, our primary dataset 

consists of the prices collected for the 26 grocery products in the ACCRA sample.13 These 

prices typically correspond to a distinct food product, such as a pound of T-Bone steak or a 2 

liter bottle of Coca-Cola, sold at a specific retail outlet on a given day. For the grocery items, 

CCER collects multiple price quotes for each item within a market at a point in time. For 

example, the surveyors collect multiple price quotes for 2 liter bottles of Coca-Cola in the first 

quarter of 2006. Smaller markets tend to have fewer price quotes per item than large markets.14 

In our data we observe the retail banner that a price quote corresponds to but not the specific 

retail location that was sampled. Thus, while we know that prices came from a Safeway store 

located in the San Francisco, CA metropolitan area in the second quarter of 2009, we do not 

 

11 See the Council’s web page for more details http://www.coli.org/. 12 As is discussed in more detail below, data collection by the individual Chambers of Commerce is voluntary. As a 

result, data collection does not occur in all time periods in all markets; that is, the dataset is not a balanced panel. In 

this study we limit our attention to those markets that have price data for at least 10 of the 14 quarters in our sample 

period. This restriction reduces the number of cities in our sample from 357 to 228. 

13 See Appendix A for a list of the grocery products and the expenditure weight assigned to each product. 14 In our data we observe markets with as few as 1 retailer surveyed within a quarter while others have more than 

thirty. In the median market/quarter 5 retail outlets and 4 retailers are surveyed; i.e., in the median market prices 

two outlets of a single retailer have been visited. 
7 

know precisely which Safeway store was sampled. As a result, in our empirical analysis we 

must treat the broad geographic region as the geographic unit of observation. 

We also use price quotes for some non-grocery items in the CCER data to control for 

unobserved market-specific retailing cost shocks, such as the local price of labor, that could 

affect grocery prices.15 We have identified four items that are unlikely to be sold at a 

supermarket that would be sold at a retailer facing similar costs as a supermarket: a men’s dress 

shirt, boy’s jeans, women’s slacks, and a three-pack of tennis balls. In contrast to the grocery 

data, we observe only a single price of these items (rather than multiple price quotes from 

different retailers). 

The CCER data is particularly well suited to our study. First, it contains prices on a 

broad set of supermarket products designed to measure the typical “market basket” of 

consumers’ food purchases. Second, the data covers more geographic regions within the U.S. 

than any other publicly accessible pricing data set we are aware of. This allows us to study many 

mergers and gives us a great deal of flexibility in identifying potential comparison cities to use in 

both our difference-in-difference analysis and in constructing a synthetic control. Third, we 

were able to collect a relatively long panel of data (5 years). 

There are two key relative weaknesses of the CCER data. The first is data quality. 

Supermarket scanner data (often provided by A.C. Nielsen or IRI) is recorded at the supermarket 

and transmitted electronically to the data vendor minimizing the chance for measurement error. 

The U.S. Bureau of Labor Statistics (BLS) collects price quotes directly from retail outlets using 

trained surveyors under a strict protocol that has been developed over time to reduce 

measurement error. In contrast, CCER’s price collection method is more informal. While 

 

15 These controls were also used by Basker and Noel (2009). 
8 

surveyors are given a detailed set of instructions to follow in collecting prices,16 CCER does not 

enforce a formal sampling scheme. Second, the products contained in the CCER sample are, by 

construction, composed of frequently purchased supermarket products. As will be discussed in 

Section IV in more detail, the prices of frequently purchased products are more likely to be 

strongly affected by changes in competition than a randomly selected grocery product. As a 

result, the price effects of mergers we estimate likely overstate (in absolute value) the true price 

effect of the merger. As a practical matter, however, we have to accept somewhat lower quality 

data to complete our study, as the CCER is the only dataset we have been able to identify that 

has such broad geographic coverage for such a wide variety of grocery products. 17

Finally, we have also obtained annual data from the Census describing the demographic 

characteristics of the geographic markets in which the firms compete. Demographic variables 

describing a region’s population, income, and racial composition were collected at the county 

level and are aggregated to match the broad geographic regions used in this study. 

III. Market Construction and Classification 

Retailers are differentiated by location, by the types and quality of items sold in their 

stores, and by the level of service they offer consumers. As a result, market definition, 

identifying the geographic region in which retailers compete and the set of firms (or products) 

that constitute a market, can be difficult. For instance, one cannot infer different retail formats 

 

16 The instruction manual given to participants can be found at: http://www.coli.org/surveyforms/colimanual.pdf

(last visited 7/17/2012). 

17 Data sets with more detailed price data have much more limited geographic coverage. The most comprehensive 

public use scanner dataset we are aware of, the IRI Marketing Dataset, contains the prices and unit sales of 

thousands of grocery products in 30 product categories for a sample of retailers, however, only provides data for 47 

markets (Bronnenberg et al. 2008).17 Another alternative data source we considered consists of the price quotes 

collected by the BLS in constructing the Consumer Price Index (CPI), see Hosken and Reiffen (2004a), Matsa 

(2011). While the CPI price data covers many more regions than the IRI Marketing Dataset (Matsa reports using 

data from 147 metropolitan areas), its coverage is still too narrow for our study. 
9 

are important substitutes to each other because the different retail formats sell some of the same 

products. Supermarkets, club stores, supercenters, convenience stores, mass-merchandisers 

(non-supercenter outlets operated by firms Target, Kmart, or Walmart), and drug stores, for 

example, all carry some food items. However, it is unlikely that all of these retail formats are 

similarly substitutable to one another. Convenience stores offer a very limited selection of food 

products in small stores at relatively high prices, while supermarkets and supercenters offer a 

broad selection of food products (including meat and produce) at relatively low prices in large 

stores. We limit our attention to the set of retail formats that are likely to significantly affect the 

pricing of supermarkets large grocery retailers that sell food and other household goods, e.g., 

cleaning products, where consumers can purchase all of their food for a week at a single retail 

location (often referred to as offering one-stop-shopping).18 This limitation results in a set of 

retailers employing three retail formats: traditional supermarkets,19 club stores,20 and 

supercenters.21 Even this limited set of formats, however, may be too broad. Club stores, while 

offering one-stop shopping, offer much more limited product selection than supermarkets or 

 

18 Recent empirical work shows that supermarkets change their prices in response to competition from supercenters 

and possibly club retailers suggesting that these retail formats compete with one another, see, e.g., Hausman and 

Liebtag (2007), Basker and Noel (2009), and Courtemanche and Carden (2011). We are unaware of empirical work 

that directly measures substitution between supermarkets, supercenters, club stores and other types of food retailers. 19 A traditional supermarket is defined as a self-service retailer selling a full line of food products (including 

grocery, meat, and produce). See the Food Marketing Institute’s Supermarket Facts available at: 

http://www.fmi.org/facts_figs/?fuseaction=superfact. 20 Club stores are high volume retailers that typically charge members an annual fee and offer consumers a limited 

selection of a broad variety of products at lower prices than other retailers. Despite their limited product offerings 

(supermarkets or supercenters typically carry between 45,000 and 140,000 items while a club store may stock less 

than 4,000 at any point in time), club stores sell food in a large number of food categories, including meat and 

produce. According to Costco’s 2009 Annual Report, 33% of Costco’s sales were of food items with 12% of total 

Costco sales being Fresh Food items (including meat, bakery, deli, and produce). Given the very large sales volume 

of club stores (the average Costco outlet has $131 million in annual revenue), a typical club store sells more food 

items in a week than a very large traditional supermarket. 

21 Supercenters are an important and rapidly growing big-box grocery retail format. Supercenters are typically 

larger than 180,000 square feet, combining both a large supermarket and a large mass-merchandiser within the same 

store. The most well-known supercenter retailer, Wal-Mart, opened its first supercenter in 1988 and is now the 

U.S.’s largest food retailer. 
10 

supercenters, and are typically not considered important substitutes to supermarkets by antitrust 

agencies in evaluating supermarket mergers.22

Identifying the relevant geographic region that determines a store’s retail prices is 

complicated because of spatial differentiation. Retailers have an incentive to exploit highly 

localized market power by charging different prices at each retail outlet in response to highly 

localized demand and competition. There are, however, significant costs of implementing highly 

localized pricing. Determining optimal prices at the store level requires retailers to invest 

significant resources in learning localized demand, and too much price variation across a 

retailer’s stores within a metro area can affect a firm’s ability to offer consumers a consistent 

price image. As a result, we observe retailers pursuing different pricing strategies within the 

broad markets they operate in. Ashenfelter et al. (2006) report that office supply retailer Staples 

changed its pricing in response to changes in competition within the broad metropolitan area in 

which its stores operated. In contrast, in its review of the merger of A&P and Pathmark, two 

large supermarket retailers with overlapping operations in the New York City and Philadelphia 

metropolitan areas, the FTC concluded that the merger would only adversely affect competition 

in two local areas within the broader New York market.23 

In principal, we would like to estimate both how mergers affect prices within the broad 

geographic market affected by a merger and by analyzing prices at the individual store location. 

Unfortunately, like previous researchers studying competition in retail food markets (Hausman 

 

22 In its investigations of supermarket mergers the FTC has typically concluded that competition among 

supermarkets is primarily limited to other supermarkets. For example, the FTC’s complaint challenging the merger 

of A&P and Pathmark in 2007 stated that: “Retail stores other than supermarkets that sell food and grocery products 

including neighborhood “mom & pop” grocery stores, convenience stores, specialty food stores, club stores, military 

commissaries, and mass merchants do not individually or collectively effectively constrain prices at supermarkets.” 23 The FTC challenged the merger of A&P/Pathmark because of concerns about localized competition in two 

regions within the greater New York metropolitan area (Staten Island and Shirley, New York). A&P/Pathmark 

satisfied the FTC’s concerns by divesting 6 stores (of the 373 located in metro New York). See 

http://www.ftc.gov/opa/2007/11/pathwork.shtm, last visited 9/7/2012. 
11 

and Liebtag (2007), Basker and Noel (2009), Huang and Stiegert (2010) ), we do not have 

sufficient data at the store level to estimate localized price effects. As a result, all of our 

empirical analysis has to be interpreted as measuring prices within the broad geographic market 

in which retailers compete. 

Market Classification 

To implement our difference-in-difference and synthetic control estimators we must 

identify those regions that experienced a significant change in market structure as the result of a 

horizontal merger (treatment markets) and those markets that experienced no significant change 

in market structure as the result of entry, exit, or horizontal mergers (comparison markets). We 

define a market as experiencing a significant change in market structure if it experiences a 

horizontal merger, entry, or exit affecting at least five percent of the retail outlets in the market. 

In our data, some markets experience a single change in market structure while others experience 

multiple changes in market structure (e.g., entry by a retailer and a merger). To facilitate 

interpretation, we only estimate the price effect of mergers for those markets whose only 

significant change in market structure resulted from a single merger during our sample period. 

We next define two sets of potential comparison markets that we use as potential controls 

in the difference-in-difference and synthetic control analysis. The first consists of markets that 

experienced no change in market structure; that is, during the sample period the market 

experienced no entry, exit, or merger of competing firms. Because all large markets in our data 

experience some change in market structure (most often the entry or exit of a small chain 

retailer), there are no large markets in this comparison group (Hanner et al. (2011). For this 

reason we consider a second set of comparison markets that consists of markets that experienced 

a di minimis amount of exit, entry, or mergers: no single entry, exit, or merger event affected 
12 

more than 2% of stores in a market. Below we provide the technical details describing exactly 

how entry, exit, and mergers are defined in the study.

We define entry as a firm beginning operations as a grocery retailer in a market with a 

new retail brand; that is, a new firm operating in a market with a new retail brand. Our definition 

of entry does not include the sale of a local brand to a new firm that continues to operate retail 

outlets in that market under the same trade name (a new firm operating an old brand).24 We also 

do not consider within market expansion – an existing retailer opening new stores of an existing 

banner in a market – to be entry.25 

We define exit as an event that causes consumers to lose access to a brand and firm in a 

market. Parallel to entry, we do not view the sale of a retail brand to another corporate parent as 

brand exit if the subsequent owner continues to operate at least one store in an affected market 

under the original retail banner. Similarly, if a firm closes some but not all of the stores 

operating under a given banner, we consider this to be within market contraction and not exit. 

In our data we observe two types of transactions that we refer to as horizontal mergers. 

The most common type of merger we observe occurs when one firm decides to exit a market by 

selling its existing operations to a current market participant. For example, in exiting the San 

Francisco, San Jose, and Fresno, California markets in 2007, Albertsons sold its stores to 

incumbent grocery retailer Save Mart Supermarkets. Save Mart then operated those store 

locations using a new name, Lucky. The second type of transaction is a traditional merger where 

an incumbent buys all of the assets of a rival. In this scenario, the acquiring firm may or may not 

 

24 Although acquisitions of this type clearly represent a change in corporate control and the entry of a new firm 

(rather than a brand) into a region, the set of products available to consumers (brand names of retailers) do not 

change as the result of the transaction. 

25 The geographic markets used in antitrust analysis are frequently more narrow than the geographic regions we have 

defined to be markets. As a result, what we define as a market expansion (e.g., a brand with operations in Los 

Angeles, California opened a store in Ventura, California) might be interpreted as market entry in an antitrust 

analysis (if Los Angeles and Ventura were separate antitrust markets). 
13 

continue to operate the acquired firm’s stores under their prior store name. Technically, we 

identify mergers using the Trade Dimensions data by we identifying all instances where an 

incumbent firm begins operating stores that had previously been operated by a rival in a given 

market. We then searched the trade press and local newspapers to confirm that this observed 

change in store ownership was the result of either a horizontal merger or acquisition. For all but 

one of the mergers we study, we have been able to identify at least one press article identifying 

the merger.26

Our dataset contains price data for 357 different geographic regions (CBSAs), however, 

only 248 of the markets meet our inclusion criterion of having at least 10 quarters of data. Of 

these 248 markets, 27 experience at least one significant horizontal merger, 42 experience at 

least one significant entry event, and 64 experience at least one significant exit event. Many of 

the markets experiencing significant entry, exit, or a merger experience multiple changes in 

market structure during our sample period, or experience a change in market structure at the 

beginning or end of our sample period. Given our identification strategy, we cannot estimate the 

price effects of a merger for markets experiencing multiple changes in market structure within 

our sample period. When we limit attention to those markets that 1) experienced only one 

significant merger, and 2) experienced mergers in either 2007 or 2008, we are left with our 

estimation sample of 14 markets experiencing horizontal mergers.27 Table 1 provides some 

information describing the markets affected by horizontal mergers, including the number of 

stores directly involved in the merger, the number of stores in the market, the number of firms 

 

26 We have been unable to find an article documenting the merger which took place in Fort Smith, Arkansas. 27 While we include club stores in the market of firms we study, we did not observe any mergers involving club 

stores that met our inclusion criteria in the markets we examined during our time period. As noted earlier, most 

medium-sized and all large geographic markets experience some entry and exit during our sample period (Hanner et 

al. (2011)). As a result, the large markets affected by mergers also experience some entry and exit. However, as 

shown in Hanner et al. the aggregate effect of this entry and exit activity is much smaller than the changes in market 

structure caused by the mergers in the large markets studied here. 
14 

operating in the market prior to treatment, a short narrative describing each transaction, and an 

rough estimate of market concentration in the broad geographic area affected by the merger.28 

There is significant heterogeneity in the size and estimated market concentration of the markets 

experiencing mergers. Our sample consists of a number of medium-sized U.S. markets, with less 

than 100 retail outlets, and some massive markets, including metropolitan New York, 

Philadelphia, and Detroit with hundreds of retail outlets. Over half of our merger sample 

consists of highly concentrated grocery markets (with estimated HHIs greater than 2,500), while 

the remaining markets are relatively unconcentrated. New York and Philadelphia, for example, 

both have HHI’s below 1,000. This variability in market concentration provides us with an 

opportunity to determine if there is a systematic relationship between market concentration and 

the price effects resulting from consummated mergers. 

IV. Empirical Model and Results 

The goal of our study is to determine how consumer prices are affected by changes in 

market structure resulting from horizontal mergers within a retail market. The major issue faced 

by any study attempting to measure the effect of a change in market structure on retail prices is 

to develop a reasonable estimate of the counterfactual change in prices had the change not 

occurred. Simply comparing the average prices in a market affected by a horizontal merger to 

prices beforehand assumes this counterfactual change is zero, and this simple time difference 

will be biased if something unrelated but concurrent in timing to the change in market structure 

 

28 We use the estimated grocery revenues of all club, supercenter, and supermarket retailers within the broad 

geographic regions affected by mergers in calculating market concentration (HHI). Our measure of market 

concentration would likely differ from that calculated by an antitrust agency in a merger investigation. An antitrust 

agency may define the product or geographic market differently and will have access to different revenue data 

(typically from subpoena responses rather than Trade Dimensions) to calculate market concentration. In particular, 

many of the geographic regions we are examining are quite large in size, possibly larger than the antitrust markets 

suggested by the Horizontal Merger Guidelines. 
15 

also affected prices. Grocery prices, particularly meat and produce, are highly volatile. An 

increase (decrease) in the price of some food items coincident with the market event being 

studied would bias a simple time difference estimator of the event’s effect on prices upward 

(downward). 

We use two methods to calculate merger price effects. The first method follows the 

literature and estimates merger price effects using a difference-in-difference estimator; that is, 

we identify merger price effects by comparing the change in prices in markets affected by 

mergers to presumably similar markets unaffected by the merger. For this approach to be valid, it 

must be the case that the change in price of the comparison markets closely approximates the 

counterfactual change in price that would have occurred in the market affected by the event had 

the event not occurred. To validate these results we examine how similar the merger and 

comparison markets are, and determine how robust the merger estimates are to changes in the 

comparison group used in estimation. 

The second method estimates price effects using the synthetic control method suggest by 

Abadie et al. (2010). This is a data-driven technique that uses information on pre-merger prices 

and demographic characteristics to construct an explicit forecast of price for the counterfactual in 

the treatment market. 

Difference-in-Difference Estimates 

We first estimate price effects using the difference-in-difference estimation in equation 

(1) below, where the (log) of retailer i’s price in market j in quarter t is regressed on a 

retailer/market specific fixed-effect ij (γ ) , a time indicator to control for idiosyncratic factors 

affecting grocery prices in all markets in a given quarter t (δ ), an indicator set equal to one in the 
16 

post-merger period for the market affected by treatment, and (in some specifications) controls for 

time-varying market specific factors (xijt) which may affect grocery pricing. 

ijt ij t ijt ijt ijt log(p ) = γ + δ +θ(Post-Event )+ βx + e (1)

Equation (1) is estimated separately for each merger event relative to the same comparison 

group, and standard errors are clustered by CBSA. 

The first issue that arises in estimating equation (1) is how to measure grocery prices. 

We measure price by constructing a price index designed to measure the overall price of 

groceries offered by retailer at a point in time. We use the price index suggested by CCER. 

CCER’s sample is constructed to correspond to a manager’s food consumption bundle. To 

approximate this bundle, CCER has constructed expenditure weights using data extracted from 

the 2006 U.S. Consumer Expenditure Survey. We use these weights (wk) to construct a price 

index for a retailer/market/quarter (pijt) as shown in equation (2) below 

26

1

*

ijt k ijkt

k

p w p



  

(2)

where pijkt is the price of product k sold by retailer i in market j in time quarter t and wk is the 

expenditure share associated with product k. We use this price index because it likely 

corresponds to the “price” consumers consider when choosing the grocery retailer to shop at in a 

time period. The literature has typically modeled the price that consumers consider in choosing a 

retailer as the price of the entire bundle of products they will purchase rather than the price of 

any single item in the bundle, see, e.g., Bliss (1988).29

 

29 We have also considered treating the unit of observation as the price quote for a specific item, at a given retailer, 

in a market in a quarter; that is, the price of a product k (a 12 ounce box of corn flakes) sold by retailer i (Safeway) 

in market j (San Francisco) in quarter t (first quarter of 2007). The primary advantage of this approach is that there 

is much more data to estimate the pricing equation. When using this price measure we estimate a variant of equation 

(1) where the fixed-effect is redefined to correspond to a specific retailer/market/product ijk (γ ). This allows the 

mean price of each items to vary by both retailer and market. In this specification we assume the event being 
17 

We next must specify the timing of the event; that is, determine when we think the 

merger could begin having an effect on grocery pricing. To some extent, we are constrained by 

our data. While we can identify the year in which a merger took place in the Trade Dimensions 

data, we have been unable to identify precisely the quarter in which all of the mergers occurred.30 

To avoid contamination bias, we have dropped data corresponding to the year in which the event 

took place, so that the pre-event and post-event periods are clearly defined. For example, if we 

observe a merger took place in 2007, we drop data from 2007 from the regression analysis and 

define 2005 and 2006 as the pre-merger period and 2008 and 2009 as the post-merger period. 

We now turn to the issue of identifying a credible comparison group. The ideal 

comparison group would consist of grocery markets experiencing demand and supply conditions 

similar to those experiencing mergers (in particular, experiencing similar pre-merger trends in 

pricing to the merger markets) that did not experience a large change in market structure. 

Markets that experienced important exit, entry, or horizontal mergers during our sample period 

are poor candidates for the comparison market, because prices in these markets may have been 

affected by changes in market structure. As discussed in the previous section, we consider two 

candidate sets of comparison markets. The first consists of 75 markets that did not experience a 

change in market structure as the result of entry, exit, or a horizontal merger during our sample 

period. The second consists of 117 markets that experienced no entry, exit or mergers that 

affected more than 2% of stores in a market. Table 2 presents demographic characteristics of the 

merger and comparison markets to examine how these markets are similar and different. The 

 

studied affects all products sold by retailers in the market proportionately by an amountθ. As a practical matter, 

both price measures yield very similar estimated price effects. Because much of the grocery pricing literature 

suggests that consumers pick a retailer based on the price of the bundle of goods they will purchase at the retailer 

(rather than the price of any single item), we only present results in this paper using the price of the bundle. 

30 For the larger mergers we have been able to identify the dates the transactions closed, e.g., A&P’s merger with 

Pathmark. For smaller events, such as the opening or closing of a store in a small CBSA, we have not been able to 

precisely identify each event date. 
18 

average market in the narrow comparison group is much smaller than the average merger market. 

The major cause of this difference is that all of the large U.S. markets experience some entry, 

exit, and or mergers by chain grocery retailers; that is, there are no major metropolitan areas in 

the narrow comparison group. When we weaken the requirement to include those markets that 

experience small levels of entry, exit, or horizontal mergers, the average market in the broader 

comparison group becomes much larger. However, because the merger sample consists of some 

of the largest U.S. metropolitan areas, the average market in the merger group is still much larger 

than the average market in the broader comparison group. 

To address the concern that some markets in the comparison group may not be well 

matched to the treatment group, we estimate two additional specifications that limit the 

comparison group to regions experiencing similar pre-merger trending in prices to the merger 

city. To implement this we estimate a pre-merger price trend for each market in our data using 

equation (3), where for each market (j) we regress retailer i’s (log) price on a retailer/market 

fixed-effect ij (γ ) and a time trend using only data from the pre-merger period.31 

ijt ij j ijt log(p ) = γ + t+ e (3) 

Unfortunately, our ability to estimate the time trend in grocery prices is limited by our relatively 

short panel. For markets experiencing mergers in 2007 and 2008, we have only five and eight 

quarters of data, respectively, to estimate a region’s pre-event trend in prices. As a result, our 

estimate of a region’s time trend can be imprecise. Then for each treatment market we compare 

that treatment market’s estimated time trend to that of each market in the comparison group, and 

 

31 Equation (3) is estimated once for each treatment market (with data restricted to that market’s pre-event period). 

The equation is estimated twice for each comparison city using data from 2005 and 2006, and 2005, 2006, and 2007. 

When comparing a treatment region to a comparison region, we use the estimate the corresponds to the same 

estimation period. For example, in constructing a comparison group for New York (which experienced a merger in 

2007) we use estimates of αj for the comparison markets estimated with data from 2005 and 2006. 
19 

in most cases, only include those markets whose estimated time trend is within 0.005 of the 

treatment markets.32 For example, prior to the merger of A&P and Pathmark in 2007, grocery 

prices in the New York CBSA were increasing at 1.025% a quarter NewYork (α  0.01025). We 

then limit New York’s comparison group to those comparison markets where the trend in 

grocery prices is within the range of 0.5% and 1.525% per quarter. 

We next formally test to determine if a given comparison market’s time trend in the premerger period is statistically different for each merger/comparison market combination. 

Specifically, we estimate equation (4) using data on pre-merger prices for all retailers in one 

treatment market and one comparison group market, and we test whether the interaction of the 

time trend with an indicator for a retailer being in the merger market is different than zero; i.e., 

whether prices are trending significantly different from one another in the merger and 

comparison market. 

ijt ij 1 2 j ijt log p =a + α t + α t*MergerCity +e (4) 

The result of these tests appear in Table 3 where we conduct two-sided t-tests that α2  0 at the 5, 

10, 15, and 20% significance levels. We see that most of the markets affected by mergers 

experience different pre-merger pricing trends than a significant fraction of the markets in the 

comparison group. As an additional specification check we limit the comparison group used in 

the difference-in-difference analysis to those markets whose price trends are not statistically 

different (at the 10% significance level) from the trend in the merger market. 

Finally we highlight an important interpretation issue of our study. Our price index may 

over-estimate the price effects associated with mergers, because the price index is 

 

32 We require that the limited comparison group contain at least five regions. One merger market’s (Albuquerque) 

pricing was trending differently enough from the regions in the comparison group that we had to increase the bounds 

to generate a control group containing at least five regions. 
20 

disproportionately composed of items that are especially sensitive to competition. Retailers 

often pursue different pricing strategies for the different consumer products they sell. For 

example, studies of grocery retailing have found that retailers are more likely to place “popular 

products,” those consumed by a large fraction of consumers and/or products that experience a 

predictable seasonal demand spike, on sale than other grocery items (MacDonald (1998), 

Chevalier et al. (2003), Hosken and Reiffen (2004b)). Retailers likely pursue this strategy 

because offering low prices on commonly and frequently purchased products (the products 

consumers are most informed about) is a cost-effective mechanism to communicate to consumers 

a store’s price level. Low prices on infrequently purchased items (about which consumers are 

relatively uninformed) are less likely to increase a retailer’s output, because consumers are 

unlikely to respond to them (Lal and Matutes (1994), Hosken and Reiffen (2004b)). Ashenfelter 

et al. (2006) report that the office supply retailer Staples pursued this pricing strategy. Staples 

categorized products into four categories which varied in the sensitivity of the product’s pricing 

to competition.33 The prices of frequently purchased items, such as copier paper or pens, were 

adjusted much more in response to changes in competition (entry or exit of a close rival) than the 

prices of less frequently purchased items (staplers). In measuring the price of the typical bundle 

of office supplies purchased by a Staples’ consumer, Ashenfelter et al. constructed a bundle of 

products that contained products in each of the four categories weighted by the relative revenue 

share of the category. 

Unfortunately, in our study, we do not know exactly how pricing varies across the items 

included in the CCER price index. We do, however, suspect that many of the products sampled 

by CCER include items that are likely to be more sensitive to the level of retail competition than 

the average product. By construction, the CCER pricing sample consists of commonly 

 

33 The categories were known as leadership, price sensitive, non-price sensitive, and invisible items. 
21 

purchased items, such as Corn Flakes, 2 liter bottles of Coca Cola, and meat items (ground beef, 

steak, and chicken), which are more likely to be offered on sale than the average supermarket 

product and are whose prices are more likely to be highly responsive to changes in competition.34 

As a result, it is likely that our price index will be more sensitive to changes in retail competition 

than an index that included all products sold by the grocery retailer (weighted appropriately by 

relative expenditures). Despite this shortcoming, our price index should correctly estimate the 

sign of the price effect of a given treatment on a retailer’s pricing, and the relative size of the our 

estimated price effects should correspond to the relative change in a retailer’s pricing; that is, 

where we estimate a large price effect, it is likely that a retailer’s prices increased more. 

Difference-in-Difference Results 

Table 4 presents the empirical results for the difference-in-difference models estimated 

for those markets experiencing horizontal mergers. Each entry in Table 4 corresponds to the 

estimated price effect (θ from equation 1) when estimating equation (1) using data from one 

merger market and some or all of the regions in the broad control group.35 The first column in 

each table corresponds to regressions estimated using only retailer/market fixed-effects and time 

indicators as controls. The second column includes variables that measure within-market 

expansion or contraction by incumbent supermarket and supercenter retailers as controls for 

other within-market changes in retail competition which may be contemporaneous with the event 

being studied.36 The third column limits the comparison group to those markets with similar pre-

 

34 See appendix A for a complete list of items contained in the price index. 

35 We have also estimated the difference-in-difference models using the more restrictive comparison group and have 

obtained very similar estimated price effects. 

36 The ratio of the number of stores opened (closed) by expanding (contracting) incumbent supermarket 

(supercenters) firms in the current year to the total number of stores operating in that market in the previous year. 
22 

event trends in pricing to the treatment group. The fourth column limits the comparison group to 

markets whose trends are not statistically different (at the 10% level) from the merger city.37 

The difference-in-difference results for mergers are consistent with the price 

concentration hypothesis. While not all mergers in highly concentrated (unconcentrated) 

markets resulted in price increases (decreases), on average, those mergers generating the largest 

price increases take place in the most concentrated markets. We find that five mergers are 

estimated to have increased consumer prices by at least 2%, and that four of these mergers took 

place in highly concentrated markets (with estimated HHIs of more than 2500). Prices decreased 

by more than 2% following five mergers, and with one exception, these mergers took place in 

much less concentrated markets than those experiencing price increases. The remaining four 

mergers resulted in little change in consumer prices. 

 Some of our estimated price effects are very large in absolute value. As we noted 

previously, many of the items in our price index are likely to be more strongly affected by 

changes in the level of retail competition than a random item. As a result, the CCER bundle may 

overestimate the overall price effect of the merger. For example, while we estimate that the price 

of the CCER bundle fell by between 10-13% in San Francisco and San Jose following the 

purchase of Albertson’s food by Save-Mart, we strongly suspect that the reduction in grocery 

prices for all grocery items would be considerably smaller. Unfortunately, we do not have 

access to information about how the prices in the CCER bundle relate to the prices of other 

(unobserved) items sold by retailers to extrapolate from our results to directly estimate how 

 

37 We have also estimated models including the prices of other retail goods sold by non-food retailers (a man’s shirt, 

women’s slacks, boy’s pants and tennis balls) as was done by Basker and Noel (2009) and find that the inclusion of 

these variables has no effect on the estimated price effects. However, because these prices are missing for some 

market/time periods, the inclusion of these variables in the estimating equation causes us to drop some time periods 

from the estimation sample. To maximize the size of the estimation sample, we do not include these variables in the 

results presented in the paper. 
23 

much overall grocery prices changed following these mergers.38 For this reason, we interpret our 

estimated price effects as being a relative measure of how much the overall price level changed 

as the result of a change in market structure. That is, we conclude that the Save-Mart/Albertson 

transaction in San Francisco and San Jose led to the relatively large price reductions, while the 

merger of A&P and Pathmark led to more modest price reductions in New York and 

Philadelphia. 

Synthetic Control Groups 

The difference-in-differences results presented in the previous section are robust to 

several regression specifications and comparison groups. This section further assesses the 

robustness of the empirical results to the choice of comparison group using the synthetic control 

group estimator developed by Abadie et al. (2010). The synthetic control method uses observed 

characteristics of geographic markets to construct a synthetic control price (defined to be a 

weighted average of a subset of the comparison group’s prices) for each treatment (merger) 

market. For example, the best comparison price for Oklahoma City, Oklahoma is the sum of 

0.20 times the price index of Providence, RI; 0.19 times the price index of Tampa, FL; 0.16 

times the price index of Paducah, KY; 0.12 times the price index of Cedar City, UT; 0.10 times 

the price index of Tuscaloosa, AL; and smaller proportions of 10 additional CBSA or CSAs. For 

a given merger market, the optimal weights corresponding to each potential control market’s 

price are determined using data on demographics and prices from the pre-merger period for each 

potential comparison market. We estimate the price effect of the merger by taking the difference 

between the observed post-merger price of the merger city and the price of the “synthetic 

control.” Our synthetic control estimator is discussed below in more detail. 

 

38 To our knowledge, only one study, Ashenfelter et al. (2006), discussed above, has access to such data. 
24 

Implementation 

 Let ݐ ൌ 1, ... , ܶ be the time periods covered by the data and let ݐெ be the period in which 

the merger of interest occurred. Define ݅ൌ1 to be the geographic market in which the merger 

occurred, and let ݅ ൌ 2, ... , ܫ be the ܫെ1 potential comparison markets. ܲ௜௧ is the observed 

average price in market i at time t, and define ܲ௜௧

~ to be the average price that would obtain if no 

merger had occurred. The relationship between ܲ௜௧ and ܲ௜௧

~ in markets ሺ1, ... , ݅, ... , ܫሻ is given by 

ܲ௜௧ ൌ ܲ௜௧ (4) ௧௜ܦ௧௜ߙ ൅ ~

where 

൜

ܦ௜௧ ൌ 1	݂݅	݅ ൌ 1	ܽ݊݀	ݐ ൐ ݐெ

݁.ݏ݅ݓݎ݄݁ݐ݋	0 ൌ ௧௜ܦ

The variable of interest, the effect of the merger on average prices, is ߙ௜௧ for periods ݐൌݐெ ൅

1, ... , ܶ. To construct an estimate of 1t , the unobserved ܲଵ௧

~ are estimated for periods following 

the merger by taking the difference between the observed average price in market 1 and a 

weighted average of the average prices in the control markets. These weights are found by 

matching the observed attributes of control markets to the merger market in the pre-merger 

period. Specifically, we estimate the set of weights ݓ ൌ ሺݓଶ, ݓଷ, ... , 	ݓூሻ that minimize the 

difference between ܲଵ௧

~ and ∑ ݓ௜ܲ௜௧

ூ ~ ௜ୀଶ for periods ݐ ൌ 1, ... , ݐெ െ 1, where ∑ ݓ௜ܲ௜௧ூ ~ ௜ୀଶ is 

specified as a function of observed market attributes.39 The weighted sum, ∑ ݓ௜ܲ௜௧

ூ ~ ௜ୀଶ , has the 

following form: 

 

 

39 Recall that we drop data from the year a merger took place. For example, in estimating the price effects of a 

merger that took place in 2007, the pre-merger period includes data from 2005 and 2006 and post-merger data from 

2008 and 2009. 
25 

௧௜ܲ௜ݓ ∑

ூ ~ ௜ୀଶ ൌ ∑ ݓ௜ܺ௜

ூ

௜ୀଶ ൅ ∑ ߝ௜௧

ூ

௜ୀଶ , (5) 

where each ݇ൈ1 ܺ௜ vector includes market-specific attributes – population, population density, 

median per capita income, percentage of population that is black, percentage of population that is 

Hispanic, percentage of population below the poverty level, and price-levels – averaged across 

time periods 1 to ݐெ െ 1, as well as the change in each of these variables from period 1 to period 

ݐெ െ 1. The ߝ௜௧ are idiosyncratic unobserved shocks to demand and or costs in market I at time 

t. The unknown parameters and weights in equation (5) are estimated by iteratively choosing the 

ݓ ൌ ሺݓଶ,...,ݓூሻ and ܸ that minimize 

ሺܺଵ െ ∑ ݓ௜ܺ௜

ூ

௜ୀଶ ሻ′ܸሺܺଵ െ ∑ ݓ௜ܺ௜

ூ

௜ୀଶ ሻ, (6) 

where ܸ is a ݇ൈ݇ symmetric positive semidefinite matrix.40 The optimal weights, ݓ ∗ൌ

ሺݓଶ

ூݓ,...,∗

∗ሻ, are then used to estimate the desired ܲଵ௧

~ and ߙଵ௧. 

We use Stata code developed by Abadie et al. (2010) to estimate the synthetic control 

model.41 Abadie et al.’s program requires that there be a single time series for the treatment 

group being analyzed. Thus, we need to aggregate the data to the level of a market/quarter from 

a market/retailer/quarter. However, we cannot simply construct a simple average of the retailers’ 

prices in a market, because not all retailers are observed in a market in every time period; that is, 

the composition of retailers observed in a market varies over time. Therefore, we construct a 

 

40 We begin each synthetic regression at three different initial V matrices. For each initial V, we employ a fully 

nested optimization routine that searches over all diagonal positive definite matrices V and weights w for the control 

that minimizes (6). Finally, we choose the control that produced the smallest value of (6) among the three starting V 

matrices. 

41 The Stata programs implementing the synthetic treatment estimator are available at: 

http://www.mit.edu/~jhainm/synthpage.html. 
26 

price index that controls for retailer/market effects. Specifically, we regress retailer i’s (log) 

price in market j at time t on a retailer/market fixed-effect (ij ) and a series of time indicators. 

We estimate these regressions at the retailer/market level. 

ijt ij jt ijt

t

log(p ) = α + δ + e

 (7) 

The time indicator   jt  from equation (7) estimates market j’s average price at time t, holding 

retailer effects constant. We use the estimated jt  as prices in the synthetic control group 

estimator.42 Abadie et al.’s STATA programs also require a balanced panel. Hence, for a given 

merger, we limit the potential set of controls to comparison markets that report prices for each 

period reported by the merger market. 

In most cases, the synthetic control appears to closely fit the treatment market in the pretreatment period. Figure 1 provides a representative plot of Oklahoma City’s actual price index 

and its synthetic control price pre- and post-merger.43 Recall that to avoid contamination bias, the 

year the merger occurred – in this case 2007 – was excluded from the empirical analyses. 

Therfore, data from 2007 is also excluded from Figure 1. As can be seen in the figure, Oklahoma 

City’s price index matches its synthetic control nearly perfectly in the pre-merger period (Q2-

2005 thru Q3-2006), and then diverges noticeably post-merger (Q1-2008 thru Q3-2009). 

Abadie et al. do not calculate conventional standard errors for the estimated effects of 

treatment. The authors argue that in aggregate studies (like ours) the most important source of 

uncertainty is not the estimated precision of the price change within a region (which is typically 

estimated with a high degree of precision) but in the uncertainty of the methodology itself. To 

 

42 Prices are all normalized relative to the first quarter of 2006   1 j,Q 2006 δ = 0 for all regions j . All included 

treatment and comparison markets report a price in the first quarter of 2006. 43 Similar figures are available for all of the markets studied in this paper upon request. 
27 

understand the importance of this uncertainty, the authors suggest that researchers conduct 

placebo studies to compare how the measured effect of treatment for the region that actually 

received treatment compares to the measured effect of treatment for those regions that (by 

assumption) did not receive treatment. We implement this methodology as follows. For every 

merger/comparison group combination, we treat each comparison region as if it was “treated” 

and calculate the average effect of treatment. This generates a distribution of up to 116 placebo 

treatment effects (one effect corresponding to each member of the comparison group). 44 We 

then rank these effects from smallest to largest and report the percentile corresponding to the 

estimated merger price effect. Table 5 presents the synthetic control estimates and the 

percentiles of the counterfactual pricing distribution generated by the placebo study in columns 3 

and 4 respectively. For example, the estimated price effect of the supermarket merger in 

Oklahoma City is 6.2%. This price effect falls in the 94th percentile of the counterfactual pricing 

distribution; that is, 94% of the estimated price effects in the placebo group are smaller than the 

price effect for Oklahoma City. 

To facilitate comparison of the synthetic control estimates to the difference-in-difference 

estimates, we have re-estimated the difference-in-difference model using the same data used in 

the synthetic control analysis (the market-level prices generated by equation 7). We also 

generate an analogous measure of where the difference-in-difference estimate falls in the 

counterfactual distribution. For each year in which a merger event can take place (2007 or 

2008), we estimate how much the price changed following that year for each comparison market 

and market that experienced a merger in that year as in equation 8 below. 

ijt ij ijt ijt log(p ) = γ +θ (Post-Event )+ e j (8) 

 

44 Some of the initial 117 comparison cities do not have complete panels of data and had to be dropped from the 

synthetic control analysis. 
28 

We then sort the estimated price effects j (θ ) from smallest to largest for the comparison group 

and record which percentile a given merger market’s estimated price effect corresponds to. 

Columns 1 and 2 of Table 5 contain the estimated price effect and the percentile of the 

counterfactual pricing distribution to which a price effect corresponds to. For example, the 

difference-in-difference model estimates the price effect of the merger in Oklahoma City 

increased price by 7%. That price effect was larger than 93% of the price changes taking place 

in the comparison group following the merger. 

The difference-in-difference estimates in Table 5 are very similar to those estimated with 

retailer/market level data (Table 4) suggesting that the data aggregation used in equation (7) does 

not result in a meaningful change in our estimated price effects. While the difference-indifference and synthetic control models estimates are not identical, they are very similar both 

qualitatively and quantitatively. In only one case (Fort Wayne, IN) is the sign of the estimated 

price effect of the merger materially impacted by the choice of estimation method, and then only 

modestly – the estimated price effect associated with the merger in Fort Wayne varies between 

zero and -1% when using the various difference-in-difference estimators and is roughly -3% 

when using the synthetic control group. The robustness of the estimated merger price effects to 

both model specification and choice of control group suggests that mergers are likely exogenous 

to the time path of prices within the market affected by the merger; that is, the failure to 

explicitly control for the potential endogeneity of mergers does not appear to affect the 

magnitude of the estimated price effects. 

It is also important to compare the estimates of precision (standard errors) from the 

difference-in-difference results (Table 4) to the percentiles of the counterfactual distribution 

(columns 3 and 4 of Table 5). The difference-in-difference estimates should be interpreted as the 
29 

change in grocery prices post-merger in the market directly affected by the merger relative to the 

mean change in price in the comparison group. As Abadie et al. note, in most aggregate studies 

these estimates tend to be very precise, and our study is no exception. Virtually all of the 

standard errors in Table 4 are less than 0.5%. From this we can conclude that the mean change 

in price in a merger market is different than the mean change in price in the comparison group 

even for small changes in price (less than 2%). However, the results from the placebo study 

show that many markets in the comparison groups also experience changes in price similar to 

those of the treatment markets. For example, roughly 19% of the comparison markets 

experience reductions in price at least as large as those experienced by Detroit (column 2 Table 

4). The results of the placebo studies for both the difference-in-difference and synthetic control 

estimates show that relatively small estimated price effects (under 2% in absolute value) are not 

uncommon in the comparison group. As a result, we cannot be confident that relatively small 

estimated price effects were caused by the merger rather than other factors. 

 

V. Conclusion

Antitrust enforcement agencies must decide how many competitors are necessary to 

maintain competition within a market. The answer to this question depends on market specific 

supply and demand factors such as the degree of product differentiation, ease of entry and 

expansion, and the model of competition that best fits the industry. By examining a relatively 

large number of mergers taking place in the same industry that occurred at roughly the same time 

we can draw some conclusions about how changes in market structure caused by a merger affect 

prices. Despite the relative ease of entry and expansion and low aggregate profit margins, we 

find evidence that horizontal mergers in the supermarket industry can result in significant 
30 

increases in consumer prices and thereby harm consumers. The mergers that result in higher 

consumer prices are largely those that we would expect, a priori, to be potentially competitively 

harmful. When market concentration increases in highly concentrated markets as the result of a 

horizontal merger, we frequently -- but not always -- observe significant increases in grocery 

prices. Our results are consistent with the broader merger retrospective literature: mergers on the 

enforcement margin are, on average, associated with price increases. 

Because the literature has focused on estimating the price effects of mergers on the 

enforcement margin, there is little empirical evidence describing how presumably benign 

mergers affect consumer prices. Our study helps fill this gap. We find that mergers in 

unconcentrated or moderately concentrated markets grocery mergers are often associated with 

reductions in consumer prices. This result supports the presumption that competitively benign 

mergers can confer significant efficiencies that are passed on to consumer in the form of lower 

prices. Overall, our study’s findings support the use of market concentration as a screen (as 

employed by the Horizontal Merger Guidelines) to aid antitrust agencies in efficiently deploying 

scarce enforcement resources. 

We also find that there is no single price effect resulting from a merger. While the 

estimated price effect of each merger we examine is robust to model specification and estimation 

technique, the estimated price effects vary dramatically across mergers, even mergers resulting in 

similar changes in market concentration. Because of this, we advocate estimating price effects 

separately by market and transaction where sufficient data is available, rather than estimating a 

single overall price effect. By examining the empirical distribution of merger price effects we 

can learn how frequently potentially problematic mergers result in increased consumer prices, or 

how frequently likely benign mergers result in lower or unchanged consumer prices. 
31 

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Ashenfelter, Orley, Daniel Hosken, and Matthew Weinberg. 2009. "Generating Evidence 

to Guide Merger Enforcement." Competition Policy International 5(1), 67-85 

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Market

Merger 

Year Merger Description

Stores

Revenue 

Share Stores

Revenue 

Share Chains Independents 

Albuquerque, NM 2007

Albertsons buys 8 Raleys stores, 6 

continue to operate; more stores in 

purchase, Raleys continued operation 

in N. Nevada and N. California. 10 0.09 8 0.06 7 14 

Detroit-Warren-Livonia, MI 2007

Kroger acquires roughly 20 Farmer 

Jack Supermarket locations from 

Great A & P Tea Co. 73 0.15 63 0.14 19 171 

Evansville, IN-KY 2008

Houchens Industries bought all 

Buehler Foods locations including 11 

stores here. 5 0.07 11 0.13 8 9 

Fort Smith, AR-OK 2007

C V Foodliner buys 7 stores from 

CVM Inc. 10 0.08 7 0.06 4 7 

Fort Wayne, IN 2007

Kroger buys 11 stores from 

SuperValu Inc. 7 0.10 13 0.15 6 11 

Fresno, CA 2007

Save Mart Super Markets buys 5 

stores from Albertsons. 24 0.36 5 0.06 11 34 

Muskogee, OK 2007

Assoc Wholesale Grocers Inc buys 

one store from Albertsons 3 0.13 1 0.08 5 3 

New Orleans-Metairie-Kenner, LA 2007

Rouse Enterprises buys 15 stores from 

Great A & P Tea Co 4 0.02 18 0.12 7 43 

New York-Northern New Jersey-Long Island, NY-NJ-PA 2007

Great A & P Tea Co buys 111 stores 

from Pathmark. 197 0.13 112 0.09 69 769 

Oklahoma City, OK 2007

Assoc Wholesale Grocers Inc buys 12 

stores from Albertsons 13 0.04 12 0.06 11 24 

Philadelphia-CamdenWilmington, PA-NJ-DE-MD 2007 Great A & P Tea Co buys 26 stores from Pathmark;. 38 0.07 26 0.05 33 96 

Table 1: Description of Mergers Studied

Aquiring Firm Aquired Firm 

Pre-Merger Firms In 

Market
Market

Merger 

Year Merger Description

Stores

Revenue 

Share Stores

Revenue 

Share Chains Independents 

Table 1: Description of Mergers Studied

Aquiring Firm Aquired Firm 

Pre-Merger Firms In 

Market

San Francisco-OaklandFremont, CA 2007 Save Mart Super Markets buys 42 stores from Albertsons. 13 0.05 42 0.11 23 73 

San Jose-Sunnyvale-Santa Clara, CA 2007

Save Mart Super Markets buys 8 

stores from Albertsons. 8 0.06 19 0.11 21 27 

Topeka, KS 2008

Kroger buys 3 stores from Assoc 

Wholesale Grocers Inc. 7 0.27 6 0.11 4 11 
Market Characteristics

Narrow 

Comparison*

Broad 

Comparison** Merger

Price Index 2.26 2.27 2.59

0.40 0.43 0.56

Price of Boy's Jeans 19.40 19.28 20.16

3.91 3.87 3.81

Price of Men's Dress Shirt 26.43 25.67 30.32

5.88 4.98 5.79

Price of Tennis Balls 2.34 2.32 2.37

0.57 0.49 0.64

Price of Women Slacks 28.74 28.61 30.83

6.47 6.88 7.67

Total Weekly Supermarket 

Revenue (thousands) 9,459 30,481 71,893

18,050 48,303 111,037

Market Concentration (HHI) 3,368 2,914 2,334

1,211 1,195 1,104

Proportional Growth of 

Incumbent Supercenter Firms 0.012 0.024 0.007

0.029 0.010 0.004

Proportional Growth of 

Incumbent Supermarket Firms 0.007 0.020 0.013

0.017 0.011 0.009

Proportional Contraction of 

Incumbent Supermarket Firms 0.009 0.022 0.030

0.020 0.012 0.024

Median Household Income 41,087 44,043 47,898

7,629 8,324 12,313

Total Population Under the Age 

of 19 84,891 283,256 779,964

158,261 496,139 1,308,433

Population 299,437 1,002,184 2,889,977

584,330 1,707,585 4,929,187

Percentage of Population in 

Poverty 0.148 0.137 0.137

0.060 0.052 0.038

Percentage of Population African 

American 0.075 0.101 0.123

0.100 0.105 0.099

Percentage of Population 

Hispanic 0.138 0.134 0.145

0.197 0.175 0.150

Number of Markets in Group 75 117 14

*The narrow comparison group contains markets that do not experience entry, exit 

or a horizontal merger during the sample period (2005-2009).

*The broad comparison group contrains markets that do not experience any one 

entry, exit, or horizontal merger that affects more than 2% of stores in a market.

Market Type

The price statistics all correspond to the premerger time period for merger markets. 

Prices come from the first year of available data (either 2005 or 2006). All other 

statistics are calculated using 2005 measures.

Table 2: Market Characteristics by Market Type Prior to Treatment
Merger Market p-value<.05 p-value<.1 p-value<.15 p-value<.2

Albuquerque 0.74 0.88 0.91 0.93

Detroit 0.02 0.05 0.08 0.11

Evansville 0.03 0.05 0.09 0.09

FortSmith 0.26 0.35 0.41 0.49

FortWayne 0.32 0.46 0.54 0.59

Fresno 0.19 0.28 0.35 0.41

Muskogee 0.02 0.03 0.07 0.08

NewOrleans 0.11 0.14 0.22 0.29

NewYork 0.10 0.18 0.27 0.36

OklahomaCity 0.26 0.35 0.39 0.46

Philadelphia 0.19 0.26 0.33 0.37

SanFrancisco 0.20 0.32 0.34 0.37

SanJose 0.13 0.21 0.27 0.32

Topeka 0.09 0.09 0.18 0.23

Table 3: Proportion of Control Cities Whose Prices Trend Pre-Merger 

Statistically Significantly Differently than the Merger City

Confidence Level of Test
Region 1 2 3 4

Albuquerque -0.0316 -0.0327 -0.0532 -0.0441

(0.00357) (0.00416) (0.0126) (0.0101)

Detroit -0.0272 -0.0273 -0.026 -0.0274

(0.00361) (0.00406) (0.00665) (0.00380)

Evansville 0.0191 0.0192 0.0188 0.0189

(0.00348) (0.00364) (0.00506) (0.00341)

Fort Smith 0.0358 0.0359 0.0356 0.0388

(0.00344) (0.00362) (0.00746) (0.00486)

Fort Wayne -0.0112 -0.0111 -0.0108 -0.0129

(0.00367) (0.00373) (0.0115) (0.00448)

Fresno 0.0421 0.0423 0.0404 0.0441

(0.00352) (0.00390) (0.00717) (0.00450)

Muskogee -0.000405 -0.000236 -0.000752 -0.000438

(0.00344) (0.00360) (0.00732) (0.00354)

New Orleans 0.03 0.0305 0.0296 0.0299

(0.00344) (0.00457) (0.00732) (0.00382)

New York -0.0182 -0.018 -0.0152 -0.0177

(0.00350) (0.00365) (0.00831) (0.00402)

Oklahoma City 0.0582 0.0573 0.0648 0.0611

(0.00345) (0.00451) (0.00785) (0.00486)

Philadelphia -0.0437 -0.0425 -0.0476 -0.0438

(0.00345) (0.00443) (0.00458) (0.00393)

San Francisco -0.133 -0.133 -0.135 -0.134

(0.00347) (0.00467) (0.00411) (0.00399)

San Jose -0.105 -0.107 -0.104 -0.105

(0.00342) (0.00542) (0.00635) (0.00378)

Topeka 0.0869 0.0874 0.0929 0.087

(0.00342) (0.00384) (0.00540) (0.00346)

Specification

Market/Retailer Fixed-Effects xxxx

Quarter Indicators xxxx

Broad Comparison Group xxxx

Measures of within market 

expansion or contraction by 

incumbent retailers.

x

Limit comparison to those with 

similar pre-merger trending x

Limit comparison group to markets 

where pre-merger trending is not 

statistically different (at 0.1 level) 

than merger region.

x

Standard errors (in parentheses) are clustered by region.

Dependent variable is the log of a retailer's price index in a region/quarter.

Table 4: Estimated Effects of Mergers on Price: Difference-in-Difference
Merger Market Pre-Merger HHI Coefficient

Percentile Of 

Counterfactual 

Distribution Coefficient

Percentile Of 

Counterfactual 

Distribution

Albuquerque 3251 -0.035 0.10 -0.026 0.27

Detroit 1260 -0.020 0.19 -0.053 0.14

Evansville 3331 0.013 0.54 0.006 0.53

Fort Smith 5278 0.048 0.84 0.065 0.94

Fort Wayne 2943 -0.001 0.56 -0.032 0.20

Fresno 1705 0.054 0.89 0.040 0.88

Muskogee 3375 0.010 0.47 -0.007 0.51

New Orleans 3462 0.035 0.75 0.019 0.75

New York 597 -0.009 0.32 -0.017 0.40

Oklahoma City 3961 0.070 0.93 0.062 0.94

Philadelphia 817 -0.035 0.11 -0.040 0.17

San Francisco 2152 -0.117 0.03 -0.115 0.04

San Jose 1729 -0.095 0.03 -0.078 0.09

Topeka 3572 0.077 0.96 0.060 0.92

Table 5: Estimated Price Effects Mergers

Comparison of Difference-in-Difference and Synthetic Control Estimates

Note: The difference-in-difference models include time indicators and market fixed-effects.

Difference-in-Difference Synthetic Control

Product Expenditure weight Product Description

T-bone Steak 0.031121 Price per pound

Ground Beef or 

Hamburger 0.031121 Price per pound, lowest price, min 80% lean

Sausage 0.03751

Price per pound, Jimmy Dean or Owens Brand, 

100% pork

Frying Chicken 0.03648 Price per pound, whole fryer

Chunk Light Tuna 0.035243 6.0 oz can, Starkist or Chicken of the Sea

Whole Milk 0.034522 Half-Gallon Carton

Eggs 0.008141 One dozen, Grade A large

margarine 0.004288 One Pound, Cubes, Blue Bonnet or Parkay

Parmesan Cheese, 

grated 0.065746 8 oz. cannister, Kraft brand

Potatoes 0.030524 10 lb., white or red

Bananas 0.056884 Price per pound

Iceberg Lettuce 0.026154 Head, approximately 1.25 pounds

Bread, White 0.08512

24 oz loaf, lowest price, or prorated 24 oz. 

equivalent, lowest price

Fresh Orange 

juice 0.016255

64 oz (1.89 liters) Tropicana or Florida Natural 

Brand

Coffee, vacuumpacked 0.03650111.5 oz. can, Maxwell House, Hillse Brothers, or 

Folgers

Sugar 0.03514 4 pound sack, cane or beet, lowest price

Corn Flakes 0.038438 18 oz., Kelloggs's or Post Toasties

Sweet Peas 0.012675 15-15.25 oz. can, Del Monte or Green Giant

Peaches 0.013836

29 ounce can , Hunts, Del Monte, Libby's, or 

Lady Alberta

Facial Tissues 0.051628 200-count box, Kleenex Brand

Dishwashing 

Powder 0.051628 75 oz. Cascade dishwashing powder

Shortening 0.017765 3 pound can, all vegetable, Crisco brand

Frozen Meal 0.099643

8 to 10 oz., frozen chicken entrée, Health Choice 

or Lean Cuisine brand

Frozen Corn 0.012675 16 oz., whole kernel, lowest price

Potato Chips 0.078015 12 oz. plain regular potato chips

Soft Drink 0.052947 2 liter Coca Cola excluding any deposit

Appendix Table 1: Items in Grocery Bundle
Americus, GA Flagstaff, AZ Odessa, TX

Ames, IA Gainesville, FL Omaha-Council Bluffs, NE-IAAtlanta-Sandy Springs-Marietta, GA Garden City, KS Orlando-Kissimmee, FLAustin-Round Rock, TX Grand Junction, CO Paducah, KY-ILBakersfield, CA Greenville-Mauldin-Easley, SC Palestine, TXBaltimore-Towson, MD Hartford-West Hartford-East Hartford, CT Phoenix-Mesa-Scottsdale, AZBeaumont-Port Arthur, TX Hays, KS Pittsburgh, PABellingham, WA Hickory-Lenoir-Morganton, NC Portland-Vancouver-Beaverton, OR-WABoise City-Nampa, ID Hot Springs, AR Prescott, AZBoston-Cambridge-Quincy, MA-NH Houston-Sugar Land-Baytown, TX Providence-New Bedford-Fall River, RI-MABradenton-Sarasota-Venice, FL Idaho Falls, ID Pueblo, COBridgeport-Stamford-Norwalk, CT Indiana, PA Punta Gorda, FLBrownsville-Harlingen, TX Indianapolis-Carmel, IN Quincy, IL-MOBurlington, IA-IL Ithaca, NY Raleigh-Cary, NCCarlsbad-Artesia, NM Jefferson City, MO Richmond, VACedar City, UT Kansas City, MO-KS Riverside-San Bernardino-Ontario, CAChampaign-Urbana, IL Kennewick-Pasco-Richland, WA Salt Lake City, UTCharleston, WV Kodiak, AK San Angelo, TXCharleston-North Charleston-Summerville, SC Lafayette, IN San Antonio, TXCharlottesville, VA Lake Charles, LA Seattle-Tacoma-Bellevue, WACincinnati-Middletown, OH-KY-IN Lake Havasu City-Kingman, AZ Sheboygan, WIColumbia, MO Lancaster, PA Shreveport-Bossier City, LAColumbus, OH Laramie, WY Springfield, MOCorpus Christi, TX Las Cruces, NM St. Cloud, MNDallas-Fort Worth-Arlington, TX Las Vegas-Paradise, NV St. George, UTDanville, IL Lima, OH St. Louis, MO-ILDayton, OH Little Rock-North Little Rock-Conway, AR Tampa-St. Petersburg-Clearwater, FLDecatur, IL Los Angeles-Long Beach-Santa Ana, CA Tucson, AZDenver-Aurora-Broomfield, CO Louisville/Jefferson County, KY-IN Tuscaloosa, ALDes Moines-West Des Moines, IA Mason City, IA Twin Falls, IDDodge City, KS McAllen-Edinburg-Mission, TX Tyler, TXDubuque, IA Memphis, TN-MS-AR Valdosta, GADurham-Chapel Hill, NC Miami-Fort Lauderdale-Pompano Beach, FL Virginia Beach-Norfolk-Newport News, VADyersburg, TN Mobile, AL Washington-Arlington-Alexandria, DC-VA-Erie, PA Morristown, TN Waterloo-Cedar Falls, IAFairbanks, AK Nashville-Davidson--Murfreesboro--Franklin, TN Wausau, WIFargo, ND-MN New Haven-Milford, CT Wilmington, NCFarmington, NM Gunnison CO Worcester, MAFindlay, OH Norwich-New London, CT Yuma, AZ

Appendix Table 2: List Of Broad Comparison Group Cities