94666 NOTE LICENSE TO SELL: THE EFFECT OF BUSINESS REGISTRATION REFORM ON ENTREPRENEURIAL ACTIVITY IN MEXICO Miriam Bruhn* Abstract—This paper estimates the economic effects of a recent reform mal businesses registering, the paper traces out the impact of this that simplified business entry regulation in Mexico. The reform was intro- increase in competition on employment, consumer prices, and incum- duced in different municipalities at different points in time. Using micro- level data, I find that the reform increased the number of registered busi- bents’ income. The results show that the fraction of wage earners in nesses by 5%. This increase was due to former wage earners’ opening eligible industries increased by 2.2%.3 In particular, people who were businesses. Former unregistered business owners were not more likely to previously not employed were more likely to work as wage earners register their business after the reform. The reform also increased wage after the reform. employment by 2.2%. Finally, the results imply that the competition from Moreover, by increasing competition, the reform benefited consu- new entrants decreased the income of incumbent businesses by 3%. mers and hurt incumbent businesses. First, the reform decreased the price level by about 1%. Second, the income of incumbent registered I. Introduction businesses declined by 3%. Finally, the income of the previously not USINESS entry regulation varies widely across the world. Djan- employed increased after the reform by about 6%. B kov et al (2002) find that the number of procedures for register- ing a business ranged from 2 in Canada to 21 in the Dominican Concurrent to this paper, Kaplan, Piedra, and Seira (2006) have inves- tigated the effect of the same business registration reform in Mexico on Republic in 1999. Given these large differences, an important ques- firm creation. A key difference is that they use registration and employ- tion to study is the effect of entry regulation on economic outcomes. ment data from the Social Security Institute (IMSS), while my paper Most work on this question has been based on cross-country studies, uses household data from the labor market survey. The IMSS data do yet such an approach suffers from identification problems, such as not capture registered firms without employees since owners do not typi- reverse causality and omitted variable bias.1 Recent work has also cally register themselves with social security.4 Moreover, not all regis- looked at cross-country, cross-industry variation for an identification tered firms with employees are in the IMSS database since a significant strategy.2 However, this approach cannot quantify the overall effect fraction of owners does not register their workers with IMSS. Conse- of differences in entry regulation, since all estimates are relative to a quently, the estimated increase in the number of registered firms in benchmark value of the ‘‘natural’’ rates of entry within industries. Kaplan et al. is 7.6 times smaller than the increase in the number of This paper uses within-country microlevel data to examine the registered firms estimated in this paper. Another difference is that I pro- effect of simplifying entry regulation on registration, employment, vide direct evidence that newly registered firms are not previously exist- prices, and income. Specifically, it exploits cross-municipality and ing informal firms, but instead new firms created by former wage earn- cross-time variation in a recent business registration reform in Mex- ers. Furthermore, Kaplan et al.’s data do not include information on ico to measure the effects of this reform, which allows for establish- income. My paper, however, identifies the effect of the reform on the ing causality more convincingly than cross-country studies do. The income of different prereform occupation groups. results show that the reform increased the number of registered busi- nesses by 5% in eligible industries, supporting the finding of the II. The Mexican Rapid Business Opening cross-country literature that less regulation leads to more entry. System Reform The use of microdata also makes it possible to trace out the effects The reform consisted in creating a Rapid Business Opening System of the reform on the functioning of the product and labor markets. (SARE) in various municipalities and was organized by the Federal Many economists have argued that barriers to entry harm consumers Commission for Improving Regulation (COFEMER). COFEMER by raising prices and thwarting employment growth. This paper first had to coordinate with municipality governments on implementing examines whether the reform led to the creation of new firms or to the the reform since many business registration procedures are set locally registration of existing informal businesses. Having shown that it led in Mexico. SARE was implemented in different municipalities at dif- to previous wage earners’ opening new businesses, rather than infor- ferent times, staring in May 2002. By September 2006, 103 municipa- lities had a SARE.5 Received for publication March 3, 2008. Revision accepted for publica- SARE was successful in simplifying local business registration tion September 29, 2009. * Development Research Group, World Bank. procedures. After reform, the averages for the number of days, proce- I am very grateful to Abhijit Banerjee, Esther Duflo, and David Autor dures, and office visits required to register a business all decreased for their advice and support. I thank Josh Angrist, Simeon Djankov, Jin significantly, falling from 30.1 to 1.4, from 7.9 to 2.7, and from 4.2 to Li, David McKenzie, Ben Olken, Filipa Sa, Olga Shurchkov, Tavneet 1, respectively. The reduction in registration procedures from 8 on Suri, Bilal Zia, and participants in the field lunches at MIT for valuable comments. I am also grateful to Roberto Villarreal from the Presidential 3 Office for Public Policy for his help during my stay in Mexico City and to This finding is in line with Bertrand and Kramarz (2002), which shows the staff of the COFEMER for taking the time to tell me about the motiva- that French regions with stricter enforcement of a zoning law have lower tion and implementation of the reform. employment growth in the retail trade industry. 1 4 For example, Loayza, Oviedo, and Serve ´ n (2005) and Djankov, According to my data, in the prereform period, 40% of registered busi- McLiesh, and Ramalho (2006) provide evidence that countries with less reg- nesses had no employees. Among these firms, less than 0.2% of owners ulation grow faster. were registered with IMSS. 2 5 Klapper, Laeven, and Rajan (2006), as well as Fisman and Sarria- There are 2,454 municipalities in Mexico, but 94% of the population Allende (2004), show that countries with heavier entry regulation have and 98% of economic activity are concentrated in 450 municipalities. lower firm entry and lower growth in value added in naturally high-entry These 450 include 99 of the 103 municipalities that had a SARE by Sep- industries. tember 2006. The Review of Economics and Statistics, February 2011, 93(1): 382–386 Ó 2011 by the President and Fellows of Harvard College and the Massachusetts Institute of Technology NOTE 383 TABLE 1.—PREREFORM AVERAGES OF OUTCOME VARIABLES Coefficient on Coefficient on Quarter of Difference in Quarter of Early Adopters’ Late Adopters’ Difference in Implementation Average Implementation Average (Levels) Average (Levels) Average Levels (Levels) Changes (Changes) (1) (2) (3) (4) (5) (6) Wage earner dummy 0.5015 0.4941 0.0074 À0.0021 À0.0125 0.0022 (0.5000) (0.5000) (0.0119) (0.0024) (0.0079) (0.0016) Low-risk wage earner dummy 0.2998 0.2869 0.0129 À0.0019 À0.0106 0.0016 (0.4582) (0.4523) (0.0170) (0.0029) (0.0091) (0.0018) High-risk wage earner dummy 0.2017 0.2072 À0.0055 À0.0002 À0.0020 0.0005 (0.4013) (0.4053) (0.0207) (0.0032) (0.0078) (0.0014) Registered business owner dummy 0.0842 0.0839 0.0003 0.0008 À0.0001 0.0005 (0.2776) (0.2772) (0.0064) (0.0010) (0.0046) (0.0007) Low-risk registered owner 0.0737 0.0734 0.0003 0.0007 0.0011 0.0001 (0.2612) (0.2608) (0.0058) (0.0010) (0.0044) (0.0008) High-risk registered owner 0.0105 0.0105 0.0000 0.0001 À0.0013 0.0004 (0.1019) (0.1019) (0.1019) (0.0002) (0.0014) (0.0003) Log monthly income 8.0686 7.8917 0.1769** À0.0298 À0.0335 0.0057 (0.7450) (0.8117) (0.0812) (0.0197) (0.0274) (0.0046) Standard errors in parentheses (clustered at municipality level). Early (late) adopters implemented the reform between May 2002 and March 2004 (April 2004 and December 2004). Data are from 2000 and 2001 ENE. Changes are changes in ENE municipality averages from 2000-IV to 2001-IV. Columns 4 and 6 present the coefficients of a separate regression for each variable (or change in each variable) on quarter of implementation. Significance levels: * 10%, ** 5%, *** 1%. average to less than 3 corresponds to going from the 30th percentile Table 1 includes summary statistics for the outcome variables in in registration procedures to the 2nd percentile, which is equivalent to the prereform period (2000-II to 2001-IV). Half of the people in my going from Bangladesh or Kazakhstan to Australia or New Zealand sample are wage earners. Another 8.4% are registered business own- (World Bank, 2006). ers. Most registered business owners are in low-risk industries SARE applied only to nongovernmental firms in industries that do (7.3%), while only 1.1% own registered high-risk businesses. not require special permits and do not present a serious risk to public health, public security, or the environment. These eligible ‘‘low-risk’’ industries made up 55% of all industries and 80% of operating firms, IV. Identification Strategy typically micro, small, or medium-size businesses—for example, com- This paper uses the cross-municipality and cross-time variation in merce and restaurants. Examples of high-risk or regulated industries are the implementation of the reform to determine its effect. The data chemical production and transportation (including taxis). cover all quarters from 2000-II to 2004-IV. I restrict the sample to the Postreform administrative data from one municipality, Guadala- 34 municipalities in my data that adopted the reform by December jara, show that most of the newly registered business were video 2004. This allows me to exploit only variation in the time of adoption, game console rental, computer rental, small grocery stores, clothing holding the decision to adopt fixed. One reason for not including stores, home-style food-to-go vendors, and beauty salons, with an municipalities that adopted after December 2004 or that have not yet average investment of US$5,471 and an average employment per firm adopted is that these municipalities become increasingly less compar- of 1.27. able to the ones that adopted the reform early.7 Moreover, the part of the analysis that looks at the effects of the reform on different prere- form groups requires me to observe individuals at least once before III. Mexican Employment Survey Data the reform was implemented. The fact that adoption of the reform varied across municipalities My main outcome data come from the Mexican National Employ- and across time makes it possible to control for municipality-specific ment Survey (ENE), the survey that the Mexican government relies and time-specific effects. The effect of the reform is thus identified on for calculating unemployment statistics and the size of the infor- off cross-municipality differences in the reform dummy over time. mal sector. The ENE has been conducted quarterly since 2000-II and The identification strategy is valid as long as the changes in outcome covers a random sample of approximately 150,000 households. Each variables over time would be similar across municipalities in the household remains in the survey for five consecutive quarters. I use absence of the reform. In particular, the identification strategy may be data for 2000-II to 2004-IV (nineteen quarters in total). After 2004- violated if the implementation of the reform followed a specific pat- IV, the ENE was changed to a new survey, and these data have not tern in terms of municipality characteristics that are related to been made publicly available. changes in outcomes. To gauge whether there was such a specific pat- I construct two of my main outcome variables by creating dummy tern of implementation, I performed a number of checks. variables for each person in the sample, indicating whether the person First, I interviewed staff members at the COFEMER who were in (a) owns a registered business or (b) is a wage earner.6 I also use charge of implementing the reform. They informed me that their goal monthly income as an outcome variable and the following individual was to bring the reform to the urban municipalities with the largest background variables as controls: age, gender, marital status, and educa- volume of economic activity. However, within this set, they did not spe- tion dummies. 7 Expanding the sample to sixty municipalities by matching late-adopter 6 Bruhn (2008) includes a detailed description of how I constructed or nonadopter municipalities to municipalities that adopted before 2004- these dummy variables. IV leads to similar results for all outcome variables. 384 THE REVIEW OF ECONOMICS AND STATISTICS TABLE 2.—IMPACT OF BUSINESS REGISTRATION REFORM Dependent Variable Low-Risk High-Risk Registered Registered Registered High-Risk Business Business Owner Business Owner Wage Earner Low-Risk Wage Wage Earner Owner Dummy Dummy Dummy Dummy Earner Dummy Dummy (1) (2) (3) (4) (5) (6) Reform dummy (SARE) 0.0031** 0.0037** À0.0006 À0.0002 0.0066** À0.0068*** (0.0015) (0.0014) (0.0005) (0.0027) (0.0029) (0.0029) R2 0.057 0.047 0.014 0.119 0.078 0.114 Number of observations 1,636,250 1,636,250 1,636,250 1,636,250 1,636,250 1,636,250 Standard errors in parentheses (clustered at municipality level). Regressions include quarter and municipality fixed effects, as well as individual-level and municipal-level control variables. Individual-level control variables are gender, age, marital status, and education dummies. Municipal-level control variables include dummies indicating whether the local party in power was the party of the president (PAN) and a dummy for whether the state and municipal ruling party were both PAN, and census variables interacted with a linear time trend. The census variables are log GDP per capita, log number of economic establishments per 1,000 capita, log fixed assets per capita, and log investment per capita from the 1999 Economic Census, converted to per capita levels using population data from the 2000 Demographic Census. Significance levels: * 10%, ** 5%, *** 1%. cify a particular pattern of implementation. In fact, they mentioned that A. Registration and Employment all local governments they approached were interested in adopting the reform, but COFEMER was not able to implement the reform in all Column 1 in table 2 shows a positive and significant impact of the municipalities simultaneously since they did not have enough personnel. reform on the number of registered businesses. Columns 2 and 3 break Second, I examine prereform changes in outcome variables. Although down the impact on registered businesses by low-risk and high-risk the identification assumption of no differential trends in absence of the businesses. Since only the low-risk businesses are eligible for the reform is fundamentally untestable, it is likely to hold if there are no reform, the increase in registered businesses should come only from initial systematic differences in trends. Column 5 of table 1 displays the low-risk businesses. This is indeed what the results in columns 2 and 3 differences in average 2000-IV to 2001-IV changes across early- and confirm. The fraction of low-risk registered businesses increased by late-adopter municipalities. Column 6 reports coefficients from a regres- 0.37 percentage points (an increase of 5% from the pre-reform level of sion of the average change in each variable on quarter of implementa- 7.4%), while there was no statistically significant change in high-risk tion. The average changes in outcome variables are not statistically dif- registered businesses. The 0.37 percentage point increase in low-risk ferent across early- and late-adopter municipalities. They are also not registered businesses corresponds to an increase in 30,678 firms for all significantly correlated with the quarter of implementation, suggesting 34 municipalities or 902 firms per municipality, on average. that the identification strategy is valid.8 Columns 5 of table 2 shows that the fraction of wage earners increased by 0.64 percentage points in low-risk industries, corre- sponding to an increase of about 2% over the prereform fraction of V. Results low-risk wage earners. Dividing the increase in wage earners (0.64) by the increase in firms (0.37) gives an average firm size of about 1.7 I obtain the main results by estimating the following regression for newly created firms. This number is much smaller than the aver- with OLS,9 age size of new firms calculated in Kaplan, Piedra, and Seira, which yict ¼ a þ bc þ ct þ dSAREct is 6.3. Since larger firms may be more likely to register their workers with IMSS, it is perhaps not surprising that they capture larger newly þ pZict þ uEC1999  t þ kPOLct þ eict ; created firms on average. Note that table 2 does not show an increase in the total fraction of wage earners. Instead, the increase in wage where the subscript i denotes individuals, c denotes municipalities, and earners in low-risk industries went along with a decrease in wage t denotes quarters. The regression includes municipality fixed effects, earners in high-risk industries, suggesting that the reform shifted bc, and quarter fixed effects, ct. The variable SAREct is the reform employment from ineligible to eligible industries. dummy, and for each municipality, it is equal to 1 for the quarter in In order to verify that the estimated effects coincide with the time which the reform was implemented and for all following quarters. Con- of the reform, I plot the coefficients of the following regression, trol variables are individual background variables, Zict, and variables X from the 1999 Economic Census (EC1999) interacted with a linear time yict ¼ a þ bc þ ct þ dl Quarterlc þ pZict þ eict ; trend, t. These variables are log GDP per capita, log number of eco- nomic establishments per 1,000 capita, log fixed assets per capita, and where Quarterlc is a set of dummy variables for lag and lead quarters log investment per capita. The regressions also include two political relative to the time of implementation in a given municipality. I limit dummies, POLct. The first one indicates whether the governing party this regression to observations that fall between QuarterÀ8 and Quar- of a municipality was the party of the president (PAN). The second terþ6 since the staggered implementation of the reform implies that I indicates whether both the municipal party and the party of the state do not observe the outcomes for all lags and leads. governor were the PAN. The standard errors of the regressions are Figure 1 shows the coefficients on the lag and lead dummies for the clustered at the municipality level. low-risk registered business owner dummy as the outcome variable. The two dashed lines are the 95% confidence bands of the estimates. 8 Bruhn (2008) also shows that there are no statistically significant dif- Between quarter À8 and quarter À1, the estimated differences tend to ferences or patterns in the changes of 1994 to 1999 Economic Census be negative or close to 0. After implementation of the reform, the dif- variables. There are some differences in the levels of 1999 Economic Census variables, which is why all the regressions control for these levels ferences become positive, reflecting the increase in registration. The interacted with a linear time trend. only exception to the positive differences after implementation is 9 Probit regressions give similar results. quarter þ5. For this lead quarter, the data set covers fewer municipa- NOTE 385 FIGURE 1.—EFFECT ON LOW-RISK REGISTRATION OVER TIME FIGURE 2.—EFFECT ON LOW-RISK WAGE EARNERS OVER TIME lities than for previous lead quarters, implying that the decrease in registration may be driven by missing data rather than by an actual TABLE 3.—IMPACT OF REFORM BY REDUCTION IN NUMBER OF PROCEDURES reversal of the effect of the reform. Figure 2 shows the effect on wage Dependent Variable: work in low-risk industries broken down by lag and lead quarters. The effect is close to 0 until the quarter of implementation. From Low-Risk Registered Low-Risk quarter 0 or quarter þ1 on, the variable is significantly and increas- Business Owner Wage Earner Dummy Dummy ingly higher than before. For 27 municipalities in my sample, COFEMER reports statistics on (1) (2) prereform and postreform registration procedures. I use this information Reform dummy (SARE) 0.0015 À0.0015 to test whether the effects are greater in municipalities with greater (0.0019) (0.0036) reductions in registration procedures. Column 1 of table 3 shows that Reform Dummy  0.0004** 0.0011*** Reduction in Procedures (0.0002) (0.0003) the reform had a bigger effect on registration in municipalities that saw R2 0.048 0.076 a higher reduction in procedures. The fraction of wage earners in low- Number of observations 1,392,477 1,392,477 risk industries also increased more in municipalities where the reduc- Standard errors in parentheses (clustered at municipality level). Regressions include quarter and muni- tion in registration procedures was greater (column 2). cipality fixed effects. They also include 1999 municipality Economic Census variables interacted with a linear time trend, as well as political party dummies and individual background variables. Significance Next, I examine whether the increase in registered businesses levels: * 10%, ** 5%, *** 1%. comes from new business creation for informal businesses registering by making use of the panel structure of the data. For each individual, I create four dummy variables specifying which occupation they held the municipality level by assigning each municipality the price index of when I first observed them in the prereform period. The four possible the city where it is located.10 CPI data exist for only 20 municipalities in occupations are registered business owner, nonregistered business my sample. As opposed to the ENE data, which have a quarterly fre- owner, wage earner, and not employed (unemployed or out of the quency, the price data come at a monthly frequency. The results in table labor force). I then drop the first period of observation for each person 5 suggest that the reform had a negative and statistically significant and use the remaining data in a regression that includes the reform effect on prices. Columns 2 and 3 break down the CPI into low-risk and dummy interacted with all four past occupation dummies. The regres- high-risk industries, showing that prices declined by about 1% in low- sion also includes quarter, municipality, and past occupation dum- risk industries. This effect may seem large, but it is in line with the pre- mies, as well as the interactions of quarter and municipality dummies vious literature. Bresnahan and Reiss (1991) find that product prices fall with occupation dummies. It also includes the individual control vari- by about 8% when going from one or two firms in the market to ables interacted with occupation dummies. between three and five firms. While the 8% estimate is for small towns Table 4 presents the regression results for this analysis. Column 1 in the United States, the effect might arguably be similar within local shows that past informal business owners are no more likely to register neighborhoods in Mexican cities. The administrative data from the their businesses after the reform. The effect on past wage earners, how- municipality of Guadalajara indicates that many of the firms that opened ever, is positive and statistically significant. Column 2 breaks down the after the reform were small neighborhood stores, which presumably effect on the fraction of wage earners in low-risk industries by prereform compete with a handful of other stores in the neighborhood. Moreover, occupation type. The results show that individuals who were previously many of the new firms were in the food and services industries, which not employed switched to being low-risk wage earners after the reform. together make up about 30% of the CPI. Moreover, columns 2 and 3 together indicate that previous wage earners Second, table 4 displays the effect of the reform on real income by moved out of high-risk industries and into low-risk industries. past occupation group. Column 5 shows a decline in real income for past registered business owners by 3%, although this effect is only B. Prices and Income 10 In my sample, only three municipalities lie in the same city (Guadala- This section first examines the effect of the reform on the Mexican jara, Zapopan, and Tlaquepaque in the City of Guadalajara). These three consumer price index (CPI), constructed by the Bank of Mexico. Since municipalities are thus assigned the same price data. All other municipali- price data are available only at the city level, I convert the price data to ties have unique observations. 386 THE REVIEW OF ECONOMICS AND STATISTICS TABLE 4.—IMPACT OF REFORM BY PREREFORM OCCUPATION Dependent Variable: Low-Risk Registered Low-Risk Wage High-Risk Wage Log Real Fourth Root of Business Owner Dummy Earner Dummy Earner Dummy Income (OLS) Real Income (1) (2) (3) (4) (5) SARE  Past Registered Owner 0.0021 0.0075 0.0030 À0.0296 À0.0431 (0.0082) (0.0058) (0.0047) (0.0201) (0.0453) SARE  Past Nonregistered Owner 0.0005 0.0021 À0.0048 0.0153 0.0808 (0.0064) (0.0082) (0.0029) (0.0236) (0.0516) SARE  Past Wage Earner 0.0025* 0.0083* À0.0148*** À0.0021 À0.0418 (0.0015) (0.0043) (0.0043) (0.0075) (0.0281) SARE  Past Not Employed À0.0004 0.0095*** 0.0014 — À0.0590* (0.0016) (0.0028) (0.0022) (0.0302) R2 0.352 0.211 0.284 0.387 0.555 Number of observations 1,051,295 1,051,295 1,051,295 419,718 733,679 Standard errors in parentheses (clustered at municipality level). Past occupation variables are prereform occupations when the person was first observed. The past not employed include the unemployed and people who are not in the labor force. The regressions in column 4 drop all individuals who were initially not employed since their past income is zero. Columns 4 and 5 include only the twenty municipalities for which the price index is available. Regressions include quarter, municipality, and past occupation dummies, as well as the interactions of quarter and municipality dummies with occupation dummies. They also include indivi- dual control variables interacted with occupation dummies, 1999 municipality Economic Census variables interacted with a linear time trend, and political party dummies. Significance levels: * 10%, ** 5%, *** 1%. TABLE 5.—IMPACT OF REFORM ON PRICES Dependent Variable: Log Consumer Price Index Log CPI for Low-Risk Industries Log CPI for High-Risk Industries (1) (2) (3) Reform dummy (SARE) À0.0067* À0.0104** À0.0047 (0.0037) (0.0045) (0.0042) 2 R 0.980 0.961 0.951 Number of observations 1,140 1,140 1,140 Standard errors in parentheses (clustered at municipality level). Regressions include month and municipality fixed effects, 1999 Economic Census variables interacted with a linear time trend, as well as political party dummies. Regressions cover 57 months for the twenty municipalities for which the price index is available. Significance levels: * 10%, ** 5%, *** 1%. marginally significant (at the 15.8% level), possibly due to the smaller REFERENCES sample size that results from using only municipalities with CPI data. Bertrand, Marianne, and Francis Kramarz, ‘‘Does Entry Regulation Hin- The results above indicate that the previously not employed were der Job Creation? Evidence from the French Retail Industry,’’ more likely to work as wage earners after the reform. These switchers Quarterly Journal of Economics 117:4 (2002), 1369–1413. should have seen an increase in their income. However, individuals Bresnahan, Timothy F., and Peter C. Reiss., ‘‘Entry and Competition in Concentrated Markets,’’ Journal of Political Economy 99:5 who were initially not employed have no income and are thus (1991), 977–1009. dropped from the log regressions in column 5. Column 6 includes the Bruhn, Miriam, ‘‘License to Sell: The Effect of Business Registration no income observations by using the quadratic root of income as the Reform on Entrepreneurial Activity in Mexico,’’ World Bank pol- outcome variable. The quadratic root mimics the logarithmic function icy research working paper no. WP4538 (2008). Djankov, Simeon, Caralee McLiesh, and Rita M. Ramalho, ‘‘Regulation well for positive numbers (see Thomas et al., 2003). The regression in and Growth,’’ Economics Letters 92:3 (2006), 395–401. column 6 shows a significant increase in the income of the previously Djankov, Simeon, Rafael La Porta, Florencio Lopez-de-Silanes, and not employed, of about 6%. Andrei Shleifer, ‘‘The Regulation of Entry,’’ Quarterly Journal of Economics 117:1 (2002), 1–37. Fisman, Raymond, and Virginia Sarria-Allende, ‘‘Regulation of Entry and VI. Conclusion the Distortion of Industrial Organization,’’ NBER working paper no. 10929 (2004). This paper uses microeconomic data to provide evidence that sim- Kaplan, David, Eduardo Piedra, and Enrique Seira, ‘‘Are Burdensome Regis- plifying entry regulation increases the number of registered busi- tration Procedures an Important Barrier on Firm Creation? Evidence nesses. It shows that after a recent business registration reform in from Mexico,’’ SIEPR discussion paper no. 06-13 (2006). Mexico, the total number of registered businesses increased by 5% in Klapper, Leora, Luc Laeven, and Raghuram Rajan, ‘‘Entry Regulation as a Barrier to Entrepreneurship,’’ Journal of Financial Economics eligible industries. This increase in registered businesses was due to 82:3 (2006), 591–629. former wage workers’ opening businesses, and not due to unregis- Loayza, Norman V., Ana Marı ´a Oviedo, and Luis Serve ´ n, ‘‘The Impact of tered business owners’ registering their businesses. Regulation of Growth and Informality: Cross-Country Evidence,’’ The paper also shows that the fraction of wage earners in eligible World Bank development policy working paper WPS3623 (2005). industries increased by 2.2% after the reform. This effect mirrors the Thomas, Duncan, Elizabeth Frankenburg, Jed Friedman, Jean-Pierre Habicht, Mohammed Akimi, Nathan Jones, Christopher McKel- cross-country results on output growth, where less complicated regu- vey, et al. ‘‘Iron Deficiency and the Well-Being of Older Adults: lation is associated with higher growth in output. Moreover, prices Early Results from a Randomized Nutrition Intervention,’’ paper decreased by 1% after the reform. presented at the Population Association of America Annual Meet- Overall, the results suggest that promoting simplification of entry ings, Minneapolis, April 2003 and the International Studies in Health and Economic Development Network meeting, San Fran- regulation is an effective policy for fostering entrepreneurial activity cisco, May 2003. and for making consumers better off by increasing employment World Bank, Doing Business in 2006: Creating Jobs (Washington, DC: opportunities and lowering prices. World Bank and International Finance Corporation, 2006).