Sophie Cottet

Luc Behaghel

Antoine Bozio

Thomas Breda

Jarkko Harju

Clément Malgouyres

David Margolis

Thomas Piketty

Payroll Tax Reductions for Minimum Wage Workers: Relative Labor Cost or Cash Windfall Effects? *

Keywords: H22, H25, H32, J23 Payroll taxes, Firm behavior, Rent sharing, Minimum wage

This paper uses administrative employer-employee data to uncover the effects of a large payroll tax reduction for minimum-wage workers in France in the 1990s. Exploiting the change in labor costs both at the job level and at the firm level, I find that the number of minimum-wage jobs increases but that these additionnal jobs stem exclusively from firms which had previously very few, or none, minimum wage workers. On the contrary, firms which already employed workers at minimum-wage levels, and thus benefit ex ante from a cash windfall, increase employment irrespective of wage levels. Overall, these results suggest that targeting cash-contrained firms, and not only groups of workers, is key for employment growth.

Introduction

The general public often perceives changes in firm taxation as borne by the firms-in particular, tax cuts are widely seen as corporate giveaways. On the other hand, economists tend to stress that taxes on firms are ultimately borne by individuals, be it workers, consumers or shareholders. Firms appear as transparent entities, which can be reduced to its production function and serve as a tax remitter and an incidence-maker.

In the case of payroll taxes, the added cost on labor that these taxes nominally impose to employers could be detrimental to minimum wage employment, as firms cannot pass it through to workers in the form of lower wages. Reducing this added cost could therefore prove beneficial to low-wage employment [START_REF] Cahuc | Baisser les charges sociales, jusqu'o ù et comment ?[END_REF]. In this perspective, many countries implemented payroll tax reductions targeting low-skilled, disadvantaged workers1 . In 2005, the OECD advocated for " [reducing] direct taxes (social security contributions and income taxes) on those with low earnings where this would shift the structure of labor demand toward low-wage workers, while protecting their incomes" [START_REF] Brandt | Assessing the OECD Jobs Strategy: Past Developments and Reforms[END_REF]. In France, payroll tax reductions for minimum wage workers have been put into place since the 1990s. Evaluations of the French payroll tax reductions have found important firm-level employment effects [START_REF] Crépon | Une nouvelle évaluation des effets des allégements de charges sociales sur les bas salaires suivi de commentaires de Yannick L'Horty et Guy Lacroix[END_REF][START_REF] Bunel | The Effects of Reduced Social Security Contributions on Employment: An Evaluation of the 2003 French Reform[END_REF], stemming from a reduction of job destruction rather than improved employment prospects from the unemployed [START_REF] Kramarz | The impact of differential payroll tax subsidies on minimum wage employment[END_REF]. They did not generate low-pay traps [START_REF] Aeberhardt | Allègements de cotisations patronales et dynamique salariale[END_REF][START_REF] Lhommeau | Les politiques d'allégements ont-elles un effet sur la mobilité salariale des travailleurs à bas salaire ?[END_REF].

Recent empirical work has however put into light unexpected effects of payroll tax cuts.

In particular, [START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF] and [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF] show that policies reducing the labor cost of a specific group of workers, irrespective of their potential effect on employment for these workers, also induce a cash windfall for firms already relying intensively on this particular type of labor. In the case of the Swedish payroll tax cut targeted at young workers, youth-intensive firms shared the rent through increased wages and also expanded more [START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF]. On the contrary, a corporate tax credit lowering labor costs for lowand medium-wage workers did not affect employment and the additional cash was shared primarily with high-skilled workers [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF]. Conversely, payroll tax rates that do not vary by labor type can affect the production process. [START_REF] Benzarti | Using Payroll Tax Variation to Unpack the Black Box of Firm-Level Production[END_REF] exploit a discrete change in Finnish payroll taxes according to firms' level of capital depreciation and find that firms facing a higher tax rate substitute away from low-skilled and manual labor.

In this paper, I analyze a major cut in payroll taxes in France which aimed at lowering the cost at the minimum wage. The 1995 and 1996 reforms reduced the minimum labor cost by 13%. These payroll tax cuts are targeted at very low wages-which is believed to be the most efficient-and with no other criteria regarding age, firm location, firm size or previous unemployment status. They provide a setting to understand the impacts of a change of the relative cost of different production factors: labor types that differ with respect to their productivity.

Using exhaustive linked employer-employee data as well as the firms' corporate tax forms, I take profit of the variation of labor cost both at the job level-the labor cost at low wage levels decreases-and at the firm level-firms which already relied on minimum wage labor see their production costs exogenously reduced. Implementing a difference-in-differences strategy at the job level, I find that the tax cut substantially increased employment at the targeted wage levels. This increase stems exclusively from firms which had no, or very few, low-wage workers before the reform. On the contrary, firms which benefit from a cash windfall due to the reform have a very steady number of workers at low wage levels and appear to slightly increase the number of jobs at higher wage levels. A difference-in-differences estimation at the firm level with respect to direct exposure to cash windfalls shows that these firms which benefited from a windfall grow faster than others.

By using a variety of lenses to examine a cut in payroll taxes, this paper sheds new light on the mechanisms through which labor costs affect firm behavior. While the distortion of the wage distribution overall shows a reaction along the relative-cost parameter, I find that this effect is dominated by a cash windfall effect. Firms that see their labor costs mechanically reduced because they already resorted intensely to minimum wage labor become suddenly less cash-constrained, and they use the additional cash to (i) hire and retain slightly more high-wage workers, (ii) increase firm-level employment overall, (iii) increase low wages above the tax reduction threshold, and (iv) increase profits more than other firms. All of the increase in the number of minimum wage workers thus comes from firms which do not benefit from a cash windfall ex ante; to profit from the payroll tax reduction, they retain more minimum wage workers than they used to and tend to trap more low-wage workers below the payroll tax cut's threshold. In other words, firms with no ex ante benefits from the reform adapt their production process and their wage policy to profit from the tax cut, while firms with ex ante benefits use the cash windfall to pursue other goals. Firms appear to be more than a black box that would mechanically repercute changes in the environment (costs, prices) into a change in its inputs, through a transparent and immutable production function.

ers that are disadvantaged on the labor market, may be more relevant in order to increase employment. Targeting minimum-wage workers succeeds in making them more employable, but giving cash to small, low-skill, liquidity-constrained firms enables them to increase employment overall. According to the policy goal with respect to unemployment, my findings show that it may be more pertinent to target firms rather than workers.

From a methodological viewpoint, these results also stress that defining an intention to treat at the firm level, based on the extent to which a firm is directly targeted by a policy-as is often done in the literature-can be misleading. Any firm is "treated" by a change in relative labor cost, and not only firms who, without any change in employment, see their costs directly affected. I show that firms which employed too few minimum wage workers to see their costs mechanically reduced by the policy account for all of the change at the low end of the wage distribution. Looking solely at firm-level effects of this direct "treatment" hides the complexity of firm's behavior in reaction to reduced labor costs. This advocates for analyzing the effects of a change of labor costs at various levels, as opposed to solely defining a "treatment" at the firm level.

Related literature. This paper contributes to several strands of literature.

First, the paper provides additional evidence of the firm-level employment effects of payroll tax cuts. I find that the more a firm benefits ex ante from the payroll tax cut, the more it increases its employment, relatively to firms with no ex ante change in their labor costs.

At the same time, firms that see their costs increase due to other concomitant changes in payroll taxes grow more slowly. These results are in line with previous studies of firmlevel effects of worker-level payroll tax cuts [START_REF] Bunel | The Effects of Reduced Social Security Contributions on Employment: An Evaluation of the 2003 French Reform[END_REF][START_REF] Crépon | Une nouvelle évaluation des effets des allégements de charges sociales sur les bas salaires suivi de commentaires de Yannick L'Horty et Guy Lacroix[END_REF][START_REF] Daunfeldt | How do firms respond to reduced labor costs? Evidence from the 2007 Swedish payroll tax reform[END_REF][START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF] and confirm that more generally, a sudden increase, or decrease, of liquidities in the firm affects firm-level employment [START_REF] Benzarti | Using Payroll Tax Variation to Unpack the Black Box of Firm-Level Production[END_REF][START_REF] Ku | Do Place-Based Tax Incentives Create Jobs?[END_REF][START_REF] Melcangi | The Marginal Propensity to Hire[END_REF]. There is little evidence on the duration of these firm-level employment effects. [START_REF] Ku | Do Place-Based Tax Incentives Create Jobs?[END_REF] find that abolishing geographically-defined payroll tax rates in Norway had lasting negative effects in regions which saw their labor costs increase. I find that in the case of the French payroll tax reduction, the employment effects are only temporary, "boosting" employment of cash-receiving firms which appear to anticipate a growth trajectory that they would have otherwise more gradually.

While I complement the evidence on firm-level effects of payroll tax cuts designed at the worker level, my results highlight the importance of diving into individual-level effects.

Most papers studying the effects of payroll taxes on employment solely define a "treatment" at the firm level, according to a measure of its exposure to the policy relying in computed ex ante total changes in costs [START_REF] Bunel | The Effects of Reduced Social Security Contributions on Employment: An Evaluation of the 2003 French Reform[END_REF][START_REF] Daunfeldt | How do firms respond to reduced labor costs? Evidence from the 2007 Swedish payroll tax reform[END_REF][START_REF] Kaunitz | Payroll Taxes and Firm Performance[END_REF][START_REF] Goos | The Impact of Payroll Tax Reductions on Employment and Wages: A Natural Experiment Using Firm Level Data[END_REF]. To the best of my knowledge, few papers combine a firm-level perspective, which allows to measure scale and cash windfall effects, with an individual-level perspective, which enables to measure the potential benefits to the targeted workers in terms of wages and employment prospects. [START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF] reveal that the payroll tax cut for young workers in Sweden did not lead to an increase in wages for eligible workers, but did improve their employment rates. [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF] establish that firms with a cash windfall due to a reduction of their corporate tax based on the amount of the wage bill of wages below 2.5 times the minimum wage did not increase their employment, and further look at the distribution of wages and at wage growth to show, using a bunching estimator, that the policy did not increase the number of "eligible" workers nor created a wage trap. My paper shows that not only is it crucial to look at the effects at different levels, but that combining these perspectives, by looking at individual effects according to the firm's exposure status, is necessary. In particular, results of this combining approach suggest that targeting low-wage workers and (indirectly) targeting cash-constrained firms leads to effects on employment of different nature: the former increases low-wage labor, while the latter boosts employment growth. In the case of the French payroll tax cut studied here, these effects coexist, but a different targeting of workers could lead to indirectly target firms which are not those with the greatest need in liquidities.

Heterogeneous effects on job dynamics according to the level of exposure of firms reveal a cash windfall channel. By bringing to light distributional effects of a cash windfall, this paper contributes to the literature examining rent-sharing behavior of firms [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF][START_REF] Howell | Do Cash Windfalls Affect Wages? Evidence from R&D Grants to Small Firms[END_REF][START_REF] Kline | Who Profits from Patents? Rent-Sharing at Innovative Firms[END_REF][START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF]. As in [START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF], the rent I examine is generated by an exogenous reduction in labor costs. Contrary to the papers in this strand of literature which look at how the rent is passed on to the average wages of workers, I investigate its effect on the distribution of wages. I highlight that firms with a positive cash windfall retain more above-threshold workers and are more prone to increase wages of below-threshold workers to the point that these workers do not give right to the payroll tax reduction anymore. I find suggestive evidence that they also hire more high-wage workers than they did before.

Finally, a payroll tax cut at the minimum wage is akin to a reduction of the minimum wage as it reduces the minimum labor cost for the employer. My paper therefore addresses the canonical question of the effect of the minimum wage on employment through a novel channel: by exploring the effects of a reduction of the minimum labor cost. My results can be interpreted as a positive effect of reducing the minimum labor cost on firm-level employment. Provided that increases and decreases of the minimum wage would have symmetric effects-which is not self-evident, partly because there could be hysteresis effects-this result is in line with other studies finding that increasing the minimum wage decreases firmlevel employment (see for instance [START_REF] Harasztosi | Who Pays for the Minimum Wage?[END_REF][START_REF] Jardim | Payroll, Revenue, and Labor Demand Effects of the Minimum Wage[END_REF]) and in contradiction with other studies finding no effect or a positive impact on employment (for instance [START_REF] Hirsch | Minimum Wage Channels of Adjustment[END_REF][START_REF] Card | Minimum Wages and Employment: A Case Study of the Fast-Food Industry in New Jersey and Pennsylvania[END_REF]. However, I argue that in this setting, the cash windfall channel could counteract or hide a "pure" minimum wage effect which would be driven by the absence or presence of frictions on the labor market.

Organization of the paper

The paper is organized as follows. Section 2 describes the nature and levels of payroll taxes in France in the 1990s and the changes induced by the 1995 and 1996 reforms. Section 3 presents the data. In section 4, I present the job-level employment effects of the payroll tax cut. Firm-level effects of the reform are presented in section 5.

Section 6 discusses those results with respect to the previous literature. Section 7 concludes.

Institutional context

Social Security Contributions

In France, payroll taxes represent 18.5% of GDP, which ranks them as the major source of public revenue. Payroll taxes encompass contributions paid by employers and employees which fund specific areas of Social Security-pensions, health, unemployment benefits and child benefits, among others-as well as less contributory sources of revenue taxed on payroll.

Payroll taxes are calculated by applying a rate to gross wages. Labor cost is the sum of gross wages and payroll taxes, and gross wages net of employee Social Security contributions constitute the net wages. The payroll tax rate for an individual worker depends on the worker's hourly gross wages: while some payroll taxes are calculated by applying a unique rate to total wage, other contributions, such as pensions and unemployment Social Security contributions, display a decreasing marginal tax rate.

In 1995, without taking into account pre-existing payroll tax cuts for low-wage workers, the average rate of payroll taxes was 38% for a minimum wage worker and for any nonexecutive employee paid less than 2.15 times the minimum wage. This rate fell to 30% for an employee paid between 2.15 and 6.45 minimum wage, 27% between 6.45 and 8.6 minimum wage, and finally to 21% for wages above. For the second and third wage groups, payroll taxes are higher if the employee is an executive.

Reforms

Payroll tax cuts targeted at low wages exist since 1993. In their first version, these tax cuts were quite modest: firms were exonerated of paying family Social Security contributions on workers paid between 1 and 1.1 minimum wage (hereafter MW), and paid half the rate of this contribution for workers with wages between 1.1 and 1.2 MW. The reduction amounted to 5.4% of the gross wage for the first group of workers, and 2.7% for the second. This meant that the effective payroll tax rate was equal to 32 and 35%, instead of the legal total rate of 38%. In July 1994, the threshold was extended to 1.3 MW (see Figure 1a). In January 1995, total exoneration of family Social Security contributions was extended to wages up to 1.2 MW (Figure 1b).

The Juppé reforms considerably increased these cuts in payroll taxes. The first reform in September 1995 implemented another payroll tax cut, representing 12.8% of gross wages at the minimum wage and continuously decreasing up to 1.2 MW (blue line in Figure 1b). The second reform in October 1996 unified both schedules to create a single un-notched tax cut with a maximum tax cut of 18.2% of gross wages and a slightly higher threshold, equal to 1.33 MW (Figure 1c). These reforms induced a large drop in low-wage labor cost. The first reform made minimum labor cost drop by an additional 10%. Labor cost for wages equal to 1.1 MW fell by 4.4%. All wages below 1.33 MW, with the exception of the [1.27, 1.3] wage range 2 , benefited from the two Juppé reforms 3 .

At the same time, other payroll tax reforms affect labor cost. Employer Social Security contributions for unemployment are slightly reduced for all workers in January 1997, from 4.18% of gross wages to 3.97%. Between 1995 and 1997, the tax rate for employers' Social Security contributions for complementary pensions increased from 3% to 3.75% for all non-executives (on the fraction of their wages below 3 "Social Security Thresholds", which in January 1995 corresponded to wages below 6.45 MW), and from 10% to 11.25% for executives on the fraction of their wages between 1 and 8 "Social Security Thresholds"-i.e., between 2.15 to 17.2 MW. 2 As you can see from Figure 1c, smoothing the payroll tax cut's schedule led to a less favorable reduction of labor cost for this small range of wages.

3 The real effect of the Juppé reforms on labor cost for those wage levels is however smaller due to two hikes in the minimum wage value in July 1995 and July 1997: these years, the minimum wage increase, which was often partly discretionary in this period, was higher than the raises in other years of the 1993-1999 period. These increases were yet well below the annual raises in the previous years.

Data

I present the administrative data that I use, and then describe the datasets constructed for the different analyses.

Administrative data sources

I combine administrative data4 with varying levels of observation, coverage and time horizons.

The primary source is the exhaustive linked employer-employee data for French firms of the private sector (DADS Postes). This data stems from the quarterly or annual declarations that firms do for the payment of Social Security contributions. These forms contain precise information on each worker in the firm: annual wages, number of hours worked and job contract period, but also occupation, executive/non-executive status, gender, age and city of residence. Firms also fill in the number of employees and the firm identifier. Each annual DADS Postes database for year N contains the job-level information for the year N -1, allowing a two-year panel at the job level. The data starts in 1993, but some variables such as the number of hours worked are poorly filled in in the first years. I use this data primarily over the period 1995-1999.

To gain more hindsight on the evolution of the distribution of jobs, I use the panel-, sampleversion of the DADS Postes, called DADS Panel, which traces back to 1967 and benefits from better data quality checks than the early years of DADS Postes. It contains information on a random sample of 1/25th of private sector workers. However, the absence or poor quality of the reported number of hours worked before 1995 imposes to restrict the sample to workers identified as working full time, in order to be able to compute their hourly wage.

Forms filled in by firms for corporate taxation contain the complete balance sheets of firms.

I use this firm-level data specifically to retrieve a measure of profits, as well as a full-year equivalent measure of the workforce5 . This data exists since 1994.

Simulation, restrictions and samples

Firm-level data. I construct firm-level data for the firm-level analysis. This data is also matched with the job-and worker-level data (see below) to ensure similar scopes of analysis.

First, I calculate the gross wage for each observation in the DADS Postes, for each year between 1994 and 1999. The employer-employee data contains annual net wages, as well as a so-called "gross" remuneration which does not correspond to the posted wage. I use the TAXIPP microsimulation model 6 to derive gross wages from the net wages, applying the employee Social Security contributions' rates and other taxes. I also calculate all components of payroll taxes and obtain the labor cost for each work contract.

Then, to get a measure of how the payroll tax reforms affect workers and firms differently, I calculate for the 1995 data, for each job, what the labor cost would have been had the 1997 payroll tax schedule apply. This counterfactual labor cost is lower than the actual labor cost for workers targeted by the payroll tax cuts, and it is higher for high-wage workers with increased complementary pension Social Security contributions .

I aggregate this comprehensive employer-employee data at the firm level, after discarding observations corresponding to hourly wages below 0.8 times the minimum wage as well as jobs which do not fall within the main Social Security scheme (régime général).

This data is merged on the firm identifier with the fiscal data. I discard firms in the financial, agricultural or energy production sectors, and firms that are not observed in both data sources.

In the unbalanced panel, I keep firms which are at least in the 1995 annual data. This is necessary to have a measure of exposure to payroll tax reductions for all firms. In a balanced panel, I keep firms observed each year of the 1994-1999 period.

In the largest sample, I keep firms with more than 2 full-year equivalent workers in 1995. This restriction limits very high growth rates for any non-negligible increase of employment in single-employee firms. In alternative sub-samples, I up this limit to 5 or 10 full-year 6 These calculations are made using the microsimulation model TAXIPP which enables precise calculations of Social Security contributions based on the information in the data, on all relevant parameters of the legislation which are detailed in the Tax and benefit tables, and on the functions associating job and firm characteristics to the amounts of Social Security contributions (and payroll taxes in particular) using the relevant parameters. Useful information in the data for Social Security contributions calculations are for the most part wages, number of hours, status (executive or non-executive), number of employees in the firm and location (Alsace-Moselle or rest of France). The microsimulation model calculates all Social Security contributions relevant for private firms, i.e., unemployment, pensions (including supplementary pensions), family, health but also all other contributions such as CSG, versement transport or contribution to the housing funds, as well as eventual reductions in payroll taxes based on wage levels (allègements généraux) or number of hours worked (Robien, Aubry I, Aubry II).

equivalents. The size of the samples are given in Table 1.

Job-level and worker-level data. I define jobs as work contracts, i.e., employee-employer pairs. I use data at the job level to analyze the changes in the distribution of wages. In the exhaustive employer-employee data, the two-year panel enables for each year N to identify new jobs (observed in N but not in N -1), destroyed jobs (observed in N -1 but not in N ) and continuing jobs (jobs observed both years).

To compensate for the lack of historical perspective, I alternatively use the DADS Panel.

The data does not contain information on the number of hours worked, so I restrict the sample to workers identified as full-time workers to adequately compute gross wages as in the comprehensive data. I do not decompose the total number of jobs per bin between new, old and continuing jobs, because restricting to full-time jobs would bias these measures and because the data features an odd hike of observations in 1994 and 1995 in the private sector, which is not reflected in the exhaustive data. I restrict the period to 1992-1999, using the data from 1991 to characterize jobs in 19927 .

For an heterogeneity analysis based on firms' characteristics, I match each of these databases with the balanced8 firm-level data.

Individual-level employment effects

In this section, I test whether the large increases of payroll tax cuts for wages around the minimum wage distorted the distribution of wages. I explore whether distortions stem from changes in hires, dismissals or retention of jobs. I present the empirical strategy in section 4.1, and the results in section 4.2. I show that the number of jobs at targeted low-wage levels increase, while remaining very stable at higher wages. Firms with different workforce composition before the reform could react differently in terms of hires, dismissals and retention, at different wage levels; hence, I test whether the wage distribution is uniformly distorted in firms with different ex ante levels of low-pay workers. Defining a measure of firms' exposure to the changes in labor costs, I find that the wage distribution evolved very differently in firms which benefited ex ante from a cash windfall thanks to the reform, and in those which did not.

Empirical strategy

I conduct an analysis at the job level to analyze the effects of the 1995-1997 reforms on the distribution of wages. In particular, I test whether, and to what extent, these changes in labor costs distorted the distribution of wages in favor of low wages. I look at whether the changes in the wages distribution stems from changes in the retention of already existing jobs, or in net created jobs. I further investigate whether continuing jobs display large wage increases.

Notations. I denote N b,t , H b,t , D b,t , C b,t and S b,t respectively the number of jobs, hires, destroyed jobs, net created jobs and continuing jobs (or stayers), in wage bin b in year t. "Hires" are jobs observed in year N which did not exist in N -1; "destroyed jobs" on the contrary are jobs observed in N -1 but not in N ; and "net created jobs" is the difference between those two aggregates. "Continuing jobs" or "stayers" are jobs that are observed in my data for two consecutive years. For destroyed jobs, the wage bin is the one observed in N -1; for all other jobs, including stayers, the wage bin is that observed in N . I further decompose between stayers who move towards a higher wage bin (of width equal to 0.1 MW) and those that don't.

I consider the ratio X b,t N b,95 for X ∈ {N, H, D, C, S}: I normalize raw numbers by the number of jobs in 1995 in the wage bin, for every bin b and year t.

Simple differences. In the baseline specification, I regress X b,t N b,95 , X ∈ {N, H, D, C, S} on wage bin dummies and their interaction with year dummies:

X b,t N b,95 = α + i γ i B i,b + k =1995 i β i,k B i,b T k,t + (1)
where

B i,b = 1 if b is bin i, and Y ear k,t = 1 if year t is equal to k. Wages range from 1 to 5
MW. The first wage bin comprises wages between 1 and 1.3 MW (treated wages), while other wage bins span 0.4 MW each. For X ∈ {N, H, C, S}, the wage bin considered is that of the wage observed in N . For X = D, the wage bin is naturally that of the wage observed in N -1.

For X = S, I present results where the wage bin is defined in N -1 when looking specifically at stayers with an upward wage mobility. This allows to study both the destination wages and the past wages of stayers. The period covered either starts in 1992 (using the panel version of the DADS, for variable N only) or in 1995. The estimated OLS βi,k is the simple difference in the normalized number of jobs X (for example, hires) in wage bin i in year k relatively to 1995.

Difference in differences. I complement this simple-difference approach by a differencein-differences strategy, whereby I define as "treated" the wages between 1 and 1.1 MW (T = 1), and as "controls" wages between 1.3 and 1.5 MW. Alternatively, I consider as treated all wages below 1.3 MW. With Y ear k,t a dummy equal to 1 if year t is equal to k, I estimate:

N b,t N b,95 = α + γ T + k =1995 δ k Y ear k,t + k =1995 β k Y ear k,t T + (2)
OLS-estimated βk gives the differential evolution of the number of jobs in treated wages between 1995 and t, compared to the control group. The β estimates for the years before 1995 provide a test of the parallel trends assumption. Note that the difference-in-differences strategy only applies for X = N , because the panel data required to have pre-1995 data imperfectly measures hires, job destruction and continuing jobs.

This strategy is similar to [START_REF] Becerra | Labor demand responses to payroll taxes in an economy with wage rigidity: Evidence from colombia[END_REF], which also estimates the effect of a targeted payroll tax cut on a normalized count measure of jobs. I introduce multiple pre-and post-reform periods.

Measuring firms' exposure to the policy. While payroll tax cuts are defined with respect to the level of wages, and therefore appear as a "treatment" at the individual level, they nonetheless affect employers differently according to the composition of their workforce.

All employers are affected by the policy, in that the Juppé reforms modify the relative cost of labor. However, these reforms also have a direct, mechanical effect on firms that already employed minimum wage workers: the additional payroll tax reductions generate an exogenous cash windfall, and the size of this additional liquidity depends on the proportion low-wage workers represent in the firm.

For each firm in the data aggregated at the firm level, I calculate the ex ante change in labor cost induced by the changes in the payroll tax schedules between 1995 and 1997. This decrease in proportions as high as 8% 9 . Most firms' variation of labor costs take negative values, but these firms are also smaller on average: they represent between 30% and 40% of total full-year equivalent employment in 1995. When grouping firms into five groups of equal total full-year employment in 1995, firms with the biggest ex ante decrease in labor costs (group Q1) has an average decrease of about 2%. Ex ante increases in labor costs are comparatively small. For each weighted quintile with positive variation of labor cost, the average value is below 0.5% (Figure 2b).

Results

Impact on total employment. Figure 3 plots the β coefficients of the simple differences regression for the total number of jobs, for a small selection of wage bins: wages between 1 and 1.3 MW ("treated" wages) and wages just above, between 1.3 and 1.7 MW, using the sampled data on the 1992-1999 period. The number of jobs in both wage bins evolve quite similarly before 1995: it slightly increases between 1994 and 1995 after having stagnated the two years before. However, their paths diverge after 1995: the number of jobs increases more at treated wages than above. The number of jobs with wages below 1.3 MW increase by 20% 10 between 1995 and 1998. By contrast, the number of jobs in the 1.3-1.7 wage bin stagnates until 1997 and very slightly increases afterwards.

The differential post-1995 trends for treated wage bins with respect to other wage bins is confirmed by the whole set of the simple differences coefficients (Figure 4a). While the total number of jobs significantly increases for wages between 1 and 1.3 MW as early as 1996, the distribution of jobs above 1.3 MW is essentially undisrupted during the whole 1992-1999 period. The post-1995 coefficients obtained using the exhaustive-but period-constrained-9 Note that a firm with only minimum wage workers would see its labor costs reduced ex ante by 9.5%. This is easily calculated, using the fact that the denominator, i.e., the labor cost with 1995 payroll taxes, is 1.326 (1.38-0.054) times the total gross wages, and that the numerator is equal to -0.128 total gross wages (difference in payroll tax cuts at the minimum wage as expressed as a percentage of gross wages). 10 Coefficients can be interpreted simply for the regressions of the total number of jobs, because the normalized number of jobs in bin b in year t is

N b,t
N b,95 . Therefore, by construction, in equation 1, α + γb is equal to 1 for every bin b, and βb,t is interpreted as an increment with respect to this value, so as a multiple of the total number of jobs in bin b in 1995.

version of the data are very similar (Figure 5).

Taking into account the small difference in trends in the numbers of jobs, the difference-indifferences estimates do confirm an increased bunching at treated wage levels. The number of jobs paid between 1 and 1.1 MW increases by 10 percentage points more than the number of jobs at higher wage levels as early as 1996, and by 10 additional percentage points in 1998 (Figure 4b). Considering as "treated" all wages below 1.3 MW, the pre-treatment coefficients are smaller in absolute values, and while the point estimates in the post-1995 period are slightly lower, the narrative remains similar (Figure A1). This increased bunching could stem from an increase in net job creation around the minimum wage and/or from higher retention of jobs at these wages. The following paragraphs investigate the sources of this evolution.

Impact on hires, job destruction and job retention.

To decompose jobs between hires, destroyed jobs and continuing jobs, I use the exhaustive version of the employer-employee data. This is done for two reasons, detailed in section 3: the restriction to full-time jobs which distorts the measure of new, destroyed and continuing jobs (for instance, a job identified as "new" in year N could have been part-time in N -1); and some odd features of the sampled data which are not mimicked by the exhaustive data. The fact that both data feature very similar results when looking at the evolution of the number of jobs post-1995 (see Figure 5) gives confidence in assuming that in the exhaustive data as well, for all wage bins, the number of jobs would have remained very similar before 1995, had the data allowed such pre-treatment analysis.

The simple difference estimates for net job creation in the 1995-1999 period show that the added bunching at the minimum wage does not stem from net job creation (Figure 6a).

While "hires" increase for the treated range of wages, job destruction increases roughly in the same proportion (Figures 6c and6d). On the contrary, the added bunching at the minimum wage appears to originate from an increase in job retention in the treated wages range (Figure 6b). The differential increase in the 1-1.3 MW wage bin between 1995 and 1999 is equal to 0.1, in units of the normalized measure of continuing jobs. This result is consistent with [START_REF] Kramarz | The impact of differential payroll tax subsidies on minimum wage employment[END_REF] which analyzes worker-level employment responses to increases and decreases in minimum labor cost in the 1990s in France. For the 1992-1999 period of minimum labor cost decreases, the authors identify workers paid in year t between the new and the t -1 minimum labor cost, i.e., workers who, at t -1, would have represented a sub-minimum labor cost for this year's legal standards, and compare their probabilities of coming from non-employment with those of workers paid at wages just above. The authors find that workers paid at the minimum wage in year t do not come more often from non-employment. This implies that any differential change in total employment at minimum wage levels with respect to slightly higher wages (which the authors do not investigate) should come from differential changes in job retention, which is indeed what I find.

The number of jobs with no significant (in terms of change of 0.1 MW-wide wage bin) upward wage mobility increases during the period at treated wage levels, but not above (Figure 7a). This could be interpreted as a wage trap: workers paid less than 1.3 MW are more likely to be "stuck" at wage levels that give right to a payroll tax reduction for their employer, in post-treatment years as compared to to 1995. Nonetheless, the probability of having an upward wage mobility also increases when this upward mobility translates into achieving a wage (still) below the 1.3 MW threshold (Figure 7c). In other words, lowering the labor cost below 1.3 MW also resulted in firms being more prone to increasing the wages of minimum wage workers-provided their new wage still gives right to a reduction in payroll taxes.

Finally, looking at the N -1 wage bin of jobs with an upward wage mobility in year N , I find that the number of jobs with N -1 wages below 1.3 MW increases, while it remains very stable at wages above (Figure 7b). This means that despite the small wage trap described above, workers with N -1 wages below the threshold have also more frequently a pay raise than before. Firms increase the number of jobs at treated wage levels (primarily through job retention), but they do not necessarily trap these added jobs in their original wage bin. Their wage can move upwards to another wage bin below 1.3 MW, as stated above, or even higher, as the number of upward-moving jobs paid between 1.3 and 1.7 MW also increases throughout the period (Figure 7c). These results confirm the worker-level analysis conducted by [START_REF] Aeberhardt | Allègements de cotisations patronales et dynamique salariale[END_REF] which showcases a higher wage mobility for workers paid below the 1.3 MW, compared to workers paid above. Additionally, these results explain the positive coefficients in the 1.3-1.7 wage bin for the most recent years, when looking at the evolution of the number of jobs (Figure 5) and of stayers in particular (Figure 6b).

Heterogeneity with respect to firms' exposure to the policy. Results so far show that, on the large population of employees of firms in the private sector, the number of jobs opening entitlement to payroll tax reductions largely increases, compared to the number of jobs at higher wages which remains roughly constant. I explore whether this increased bunching at the minimum wage is observed in all firms, or differs according to the direct, or indirect, exposure of firms to the change in payroll tax cuts.

For better statistical power, the heterogeneity analysis for the simple differences specificationincluding the regressions with the total number of jobs as the dependent variable-is based on the exhaustive employer-employee data, meaning that the regressions are run on the 1995-1999 period, thereby preventing a formal test of pre-trends. However, as for the analysis ran on the whole sample, the sign and order of magnitude of the post-1995 coefficients for the total number of jobs per bin are in line with those obtained on the panel, sample, long-run version of the data.

The results of the simple differences regression on the total number of jobs per bin, estimated separately for each of the five weighted quintiles of firms, are striking. All sub-samples display an increased bunching at the minimum wage except for the sample of jobs in firms in the first quintile, i.e., firms with negative ex ante variation of labor costs (Figures 8a to 8e).

The distribution of wages for firms in Q1 is very stable at low wage levels, with signs of a slight increase of the number of jobs at higher wages. In all other firm groups, the number of low-wage jobs expands and the number of jobs at higher wage levels remains very stable.

This additional bunching at the minimum wage increases with the level of ex ante change in labor costs. Moreover, the bigger the change in firms' labor costs, the earlier the bunching increases. In firms of Q5, the number of jobs paid between 1 and 1.2 MW increased by 50% between 1995 and 1996. By 1999, this number is 1.5 times its value in 1995.

This pattern for the total number of jobs emerges from three facts. First, in Q1, the stability of the number of low-wage jobs is not due to the compensation of hires by job destruction, or by a decrease in the number of continuing jobs balanced by an increase in net created jobs: neither hires nor job destruction or retention of workers change in these firms (Figure 9). In firms of Q2 and Q3, hires and job destruction slightly increase but in total, there is no additional net created jobs. The additional bunching at low wages stems exclusively from an increase of the number of continuing jobs (Figures 10 and11). In firms of Q4 and Q5, however, all flows contribute to the increase in the number of jobs below 1.2 MW: new jobs increase more than the destruction of jobs, and retention increases (Figures 12 and13).

The pattern of upwards wage mobility also differs. In Q1, the number of continuing jobs with upwards wage mobility with a "destination" wage above the 1.3 MW threshold increases as early as 1996, while it remains very stable in all other groups of firms (Figure 14).

Conversely, in firms without a cash windfall, and especially in Q4 and Q5, the number of continuing jobs with increasing wages with a destination wage below the 1.3 MW threshold increases. Overall, firms in Q1 appear to use the cash windfall partly to increase wages of workers, while other firms seem to increase wages only if the destination wages still give right to a reduction in payroll taxes.

Overall, firms with a cash windfall appear to react very differently to the change in relative labor cost than firms with no change, or a positive change, in labor costs ex ante. In total, they do not hire, destroy jobs or retain low-wage workers differently in the 1996-1999 period than they did in 1995. However, there appears to be slightly more hires and continuing jobs at higher wages, although estimated coefficients are not always significantly different from zero. On the other hand, in firms which did not benefit from a cash windfall, the change in the distribution of jobs seems to reflect the decrease in the relative labor cost of low-wage workers, with increased retention of these workers, combined with a wage mobility trap, and, for firms in Q4 and Q5, more net job creation.

Firm-level effects

Results at the job level show that while firms as a whole appear to be taking advantage of the change in the relative cost of labor, the wage distribution of jobs in firms directly benefiting from the reform seems roughly unchanged. While the number of hires, job destruction and job retention at the minimum wage level is strictly null, there seems to be a positive, yet not statistically significant, effect on hires and continuing jobs at higher wage levels, which indicates that benefiting from a cash windfall could participate in reshaping the firms' production function. At the firm level, benefiting from an exogenous reduction in production costs could matter for growth if firms are cash constrained. In this section, I explore whether the tax windfall affects firm-level overall employment decision and profits growth-in other words, how firms use the additional cash.

Empirical strategy

Based on the definition of firm-level direct exposure to changes in labor costs using ex ante variation of labor costs (Section 4.1), I implement a difference-in-differencse approach with multiple groups and multiple time periods. I regress the growth rate of a firm-level outcome with respect to 1995 levels on an interaction of time and exposure group dummies, taking as the time reference the pre-treatment year (1995), and as the reference group the fourth weighted decile, D4, which is the first group of firms having zero ex ante variation in labor cost (Figure 2c). This discrete measure of firm exposure allows to explore non-linearities of the treatment effect-as in [START_REF] Bunel | The Effects of Reduced Social Security Contributions on Employment: An Evaluation of the 2003 French Reform[END_REF] or [START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF] for instance-and in particular to distinguish firms with a negative change in labor costs from firms with a small increase ex ante.

Denoting Y i,t the outcome for firm i in year t, D j,i a dummy equal to 1 if firm i belongs to decile j of ex ante variation of labor cost, T k,t a dummy equal to 1 if year t = k, and E s,i a dummy for sector s, the baseline specification is:

Y i,t = α i + γ t + 10 j=1 j =3 1999 k=1994 k =1995 β j,k × D j,i × T k,t + s 1999 k=1994 k =1995 δ s,k × E s,i × T k,t + i,t (4)
α and γ are the firm and year fixed effects. δ is a sector × year fixed effect to account for potentially different economic trends. If well identified, the coefficient βj,k is the effect of being in exposure decile j-compared to being in D4-in year k on the outcome. In the reduced form estimations, Y i,t is the growth rate of employment or profits with respect to the pre-treatment year.

Using treatment groups, rather than a continuous measure of treatment (unlike [START_REF] Crépon | Une nouvelle évaluation des effets des allégements de charges sociales sur les bas salaires suivi de commentaires de Yannick L'Horty et Guy Lacroix[END_REF], allows non-linear effects of changes in the labor cost. In particular, it allows the firm-level effects of an ex ante increase in labor costs to be different from that of an ex ante reduction.11

The dummies for each year of the 1996-1999 period enables to distinguish short-term and long-term effects. Coefficients associated to the year 1994 provide a direct test of the parallel trends assumption for the 1994-1995 period.

Alternative specifications. To account for possible different growth trends for firms with different initial workforce sizes, especially as firms in the first deciles are also smaller, I control for an interaction of initial size group (ranking firms into size quintiles according to their size in 1995) and year. I also estimate equation ( 4) separately on a sample of firms with an initial size higher than 5 full-time equivalent workforce, and on a sample of firms above 10 full-time equivalent workforce in 1995, to test whether the results are driven by very small firms.

Results

First stage. Figure 15a plots the estimated β coefficients for the growth rate of the share of social charges in total labor costs. Social charges include payroll taxes and any employer contribution, such as obligatory funding for the firms' works council. This ratio grows less for firms with ex ante decreases of labor costs, and more for firms with positive ex ante variation of labor costs. In other words, ex ante variation in labor costs due to the change in payroll tax schedules is a predictor of variation in the burden of social charges. Firms that mechanically benefit from payroll tax cuts still benefit more from these reductions than other firms, meaning that the workforce composition does not change radically.

Impact of the cash windfall on employment. Figure 15b shows a clear correlation between the "exposure" decile firms belong to and their average growth rate of employment.

Between 1995 and 1997, full-year equivalent employment increases more in firms which directly benefited from payroll tax cuts, and less in firms with mechanical increase of labor costs due to the uncapping of Social Security contributions at very high wages. The cash windfall, and conversely the exogenous increase in labor costs, seem to matter for firm growth. However, this effect appears temporary as employment growth rates converge in 1998: between 1995 and 1998, firms in the first seven deciles of ex ante variation of labor costs experience overall very similar growth rates. This could be due partly to the diminishing difference in ex ante labor costs evolution between deciles in 1998, as payroll tax cuts, which benefit more to firms in D1-D3, are slightly reduced. However, the gap between 1995-1997 and 1995-1998 growth rates appears too big to be only explained by this slight change in ex ante variation of labor costs: the 1998 coefficients in Figure 15a for D1 and D2 are less negative than the 1997 coefficients, but as expected, they are not null, indicating that the change in payroll tax cuts in 1998 is small and that there is no radical change in workforce composition between 1997 and 1998. Overall, the cash windfall induced by the payroll tax cut seems to have boosted firm growth but did not increase it in the long term, because of subsequent changes in payroll tax schedules and/or because the cash windfall enables to anticipate firm growth, rather than trigger it in the long term.

Those results on firm-level employment may appear at odds with the evidence that there is no hike in low-wage jobs in firms with a strong cash windfall and the presence of such a hike in other group of firms. However, employment at higher wage levels does increase in firms in Q1 (Figure 8a). Moreover, the firm-level employment outcome in Figure 15b is calculated as the growth rate of full-year equivalent employment in the firm. As firms in groups D1 to D3 are also smaller on average, their growth rates are higher for a similar absolute increase in labor.

Impact on profits.

A cash windfall, as well as an exogenous increase in labor costs, may be passed through, or absorbed in, profits. A possible effect on product prices of the change in labor costs may also affect revenue, through an adjustment of the demand on the product market. Figure 15c plots the coefficients for the estimation of equation ( 4) for the growth rate of profits. I find that profits increase temporarily more for firms in D1 (the positive effect disappears as early as 1997), and that profits growth is slower for firms with the highest increases in payroll taxes.

Robustness

Accounting for differences in initial firm size. These results could stem from the fact that firms in the D1-D3 weighted deciles are smaller than firms in more central groups. In particular, it has been shown that smaller firms display higher employment growth [START_REF] Moscarini | The Contribution of Large and Small Employers to Job Creation in Times of High and Low Unemployment[END_REF] 12 .

I adress this concern by first adding size-year fixed effects in equation ( 4), defining "size" as the firm size quintile group in 1995. I find that the higher employment growth for D1-D3 firms with respect to D4 is less strong when accounting for differential firm growth according to initial firm size (Figure A3b): part of the higher growth for D1-D3 indeed comes from the fact that these firms are on average smaller, however there remains an effect attributable to benefiting ex ante from a cash windfall. In the same way, the relative employment growth of D8-D10 firms appears to be overestimated (i.e., the relative decrease is underestimated)

when the effect of initial firm size is not controlled for, as the firms in those groups are also smaller in average than firms in more "central" exposure groups. Effects of ex ante cash windfall on profits are robust to controlling for firm size (Figure A3c).

I also test whether results are robust to restricting the sample to firms which were bigger in 1995. Indeed, whether or not we take into account size-specific growth over the years, very small firms (with less than 5 or 10 full-year equivalents, for instance) could react differently from other, less small firms. Note that on subsamples of firms of more than 5 or 10 fullyear equivalent employment in 1995, firms with negative ex ante variation of labor costs still represent 30 to 40% of full-year equivalent employment-the reference group, with a zerovariation of labor costs on average, is still D4. Employment-growth effects are largely robust to these alternative samples of firms (Figures A4b andA5b). Restricting the sample to firms of more than 10 full-year equivalent employment in 1995 eliminates any differential growth according to initial size: adding size × year fixed effects does not change the results (Figure A6b). Moreover, the negative effect for D1 firms in 1998 disappears with the 10+ sample,

showing that it comes from very small firms (Figures A5b andA6b). The positive effect on profits of being in D1 also appears to be driven by small firms: it is robust to using the sample of firms of at least 5 full-year equivalent employment in 1995, but disappears when considering only firms of initial size greater than 10 (Figures A4c andA5c).

Unbalanced panel of firms. Facing ex ante increases or decreases of payroll taxes may correlate with the probability of a firm to exit the market. Figure A7a shows that firms from D1 and D2 face higher risks of attrition, including before 1995, which suggests that it is their fixed characteristics, rather than the cash windfall, which generates this higher attrition. This is supported by the fact that controlling for the initial size group of the firms weakens these estimates (Figure A7b). The causal interpretation of the remaining differential attrition for D1 is hazardous. In any event, this differential attrition warns against the robustness of our estimates on the balanced panel, as surviving firms are likely to be selected, and in particular growing faster. The positive coefficients for D1-D3 in 1996 and 1997 are indeed weakened when keeping all firms regardless of their being observed continuously in the data, and attributing a growth rate of -100% in year y for firms absent of the data in y (Figure A7c). The slightly positive effect on profit for D1 in 1996 disappears when using an unbalanced panel of firms (Figure A7d).

Discussion

I find that a large payroll tax cut targeted at the minimum wage succeeded in globally increasing the number of jobs at the low end of the distribution, but that all of this increase stems from firms which had too few minimum wage workers previously to be entitled to a cash windfall from this policy. Minimum wage employment in firms with a substantial ex ante cash windfall remains extremely stable, while overall employment in those firms grows more than in other firms. As these results may appear surprising, both individually and taken together, I discuss how they compare to previous literature and what they suggest of firm labor demand behavior.

Effects of the minimum wage on employment

Most work on the effects of the minimum wage on employment has either focused on particular populations, like teenagers, fast-food restaurants (as in the seminal work [START_REF] Card | Minimum Wages and Employment: A Case Study of the Fast-Food Industry in New Jersey and Pennsylvania[END_REF] or the retail industry. However, focusing on populations or industries assumed as "low-wage" is suspected of producing downward-biased estimates and raises concerns about the representativeness of these populations or sectors. More recent papers compare firms with different "bites" of the minimum wage [START_REF] Harasztosi | Who Pays for the Minimum Wage?[END_REF][START_REF] Hirsch | Minimum Wage Channels of Adjustment[END_REF][START_REF] Huang | Minimum wages and firm employment: evidence from China[END_REF][START_REF] Jardim | Payroll, Revenue, and Labor Demand Effects of the Minimum Wage[END_REF]). Yet the policy debate around the minimum wage is in general more focused on "the number of jobs destroyed" than on the effects on firms-including firm-level employment.

To my knowledge, only two recent papers approach the minimum wage question through this angle: [START_REF] Jardim | Minimum Wage Increases, Wages, and Low-Wage Employment: Evidence from Seattle[END_REF] and [START_REF] Cengiz | The Effect of Minimum Wages on Low-Wage Jobs[END_REF]. They investigate the effects of changes in the minimum wage on the number of low-pay jobs [START_REF] Jardim | Minimum Wage Increases, Wages, and Low-Wage Employment: Evidence from Seattle[END_REF] or on the wage distribution as a whole [START_REF] Cengiz | The Effect of Minimum Wages on Low-Wage Jobs[END_REF] in a similar fashion to my job-level analysis of the payroll tax cut. In particular, [START_REF] Cengiz | The Effect of Minimum Wages on Low-Wage Jobs[END_REF] apply a bunching estimator to a differences in differences strategy using a large set of US State-level minimum wage increases and find a zero net effect on employment. Indeed, they find that the destruction of below-minimum wage jobs is compensated by a hike in the bunching of the distribution at the new minimum wage. In this paper, I find that lowering the minimum labor cost has overall a positive effect on the number of jobs at the minimum wage. These results are however not contradictory:

raising the minimum labor cost can increase the number of jobs at the minimum wage if a large number of workers with productivities below the new minimum wage were already employed, and decreasing the minimum labor cost allows for a higher number of individuals with productivities below the old minimum labor cost to be employed. Moreover, in the setting I study, the pre-reform minimum labor cost was high as a ratio to median labor cost, and both [START_REF] Cengiz | The Effect of Minimum Wages on Low-Wage Jobs[END_REF] and [START_REF] Jardim | Minimum Wage Increases, Wages, and Low-Wage Employment: Evidence from Seattle[END_REF] find that job destruction increases with the level of pre-reform minimum-to-median wage ratio. Symmetrically, we could expect job creation to increase with the pre-reform bite of the minimum wage, when minimum labor cost is reduced.

Related to the minimum wage literature, my results also highlight the drawbacks of identifying minimum-wage employment effects by comparing firm-level employment of firms with more or less low-wage workers prior to the reform. When using firm-level bite as a measure of treatment, papers identify the effect of an increase of low-wage labor cost on firm behavior rather than on employment as a whole. Indeed, while I find that firms with more minimum wage workers increase firm-level employment when the minimum labor cost decreases, my paper confirms the potential worries that "non-treated" firms with no or little ex ante change in labor costs due to changes in the minimum labor cost may change their labor composition, due to the distortion of the relative cost of labor.

Moreover, at the firm level, employment effects may not be interpreted as (only) effects of the minimum wage: the cash windfall effect may confound a "pure" minimum wage effect which would be driven by the absence or presence of frictions on the labor market. Besides, the negative firm-level employment effects of ex ante increases of labor costs mimic the negative firm-level employment effects of increases of the minimum wage (see for instance [START_REF] Harasztosi | Who Pays for the Minimum Wage?[END_REF][START_REF] Jardim | Payroll, Revenue, and Labor Demand Effects of the Minimum Wage[END_REF], which stresses that the wage level concerned by a change in labor cost may not matter. In reality, symmetrically to my findings of a positive effect of a high cash windfall on profits growth, previous firm-level studies of minimum wage effects highlight a liquidity channel, which appears to be fundamental to understand any firm-level employment effect: [START_REF] Draca | Minimum Wages and Firm Profitability[END_REF][START_REF] Hirsch | Minimum Wage Channels of Adjustment[END_REF] and [START_REF] Harasztosi | Who Pays for the Minimum Wage?[END_REF] all show that firms' profits decrease with the firms' exposure to minimum wage hikes.

How do firms react to a cash windfall?

My results reveal a cash windfall mechanism: the targeted reduction in payroll taxes affects employment differently according to the amount of cash windfall, or cash loss, of firms.

Recent papers address the rent-sharing behavior of firms receiving unexpected cash. They highlight the effects on the average wages of employees, showing that part of the windfall is redistributed through higher wages to all, or a fraction, of employees [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF][START_REF] Howell | Do Cash Windfalls Affect Wages? Evidence from R&D Grants to Small Firms[END_REF][START_REF] Kline | Who Profits from Patents? Rent-Sharing at Innovative Firms[END_REF][START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF]. I take a different approach:

in the case of these targeted payroll tax reductions, firms which benefit from a cash windfall and firms that don't largely differ in terms of workforce composition, in pre-treatment year but also likely in trends, such that average wage evolutions are not comparable between groups. Rather, I investigate whether at a broader level (large groups of firms defined with respect to policy exposure), benefiting or not from a cash windfall affects the distribution of wages-bearing in mind that the policy distorts the relative cost of labor. I find that in firms benefiting from a cash windfall, overall, the wage distribution does not react at all to the change in relative labor cost. If anything, the number of jobs at higher wage levels increase.

This surprising result reminisces [START_REF] Daunfeldt | How do firms respond to reduced labor costs? Evidence from the 2007 Swedish payroll tax reform[END_REF]: the authors evaluate the firm-level effects of the payroll tax cut targeting the young in Sweden and find that firms benefiting from large labor cost savings due to this tax reduction slightly increased employment of older individuals. Contrary to my findings, whereby the number of low-wage jobs does not increase in firms with the highest cash windfall, they find that these effects on the nontargeted population are however much smaller than those on the targeted (young) individuals. Other studies find "spillover" effects of exogenous labor cost reductions, but these spillovers affect the workers' wages rather than their employment prospects [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF][START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF].

Looking at how firm-level employment changes with firms' exposure to labor cost changes, I find that firms with a cash windfall grow faster than firms with virtual no ex ante change in costs, and firms facing a cost increase due to the uncapping of Social Security contributions on high wages grow slower. This first set of results confirms what has been shown by the literature tackling firm-level employment effects of changes in payroll taxes. [START_REF] Crépon | Une nouvelle évaluation des effets des allégements de charges sociales sur les bas salaires suivi de commentaires de Yannick L'Horty et Guy Lacroix[END_REF] find a strong firm-level employment effect of the Juppé reforms. Although the overall sign matches my results, I find that defining firm exposure solely with respect to changes in payroll tax reductions (and not overall payroll tax changes) produces biased estimates, by confounding the effects of payroll tax cuts on low-wage jobs and of payroll tax increases on high wages (see Figure A2). I also nuance their results by showing that the firmlevel effects of the payroll tax cut are short term. [START_REF] Bunel | Les effets des allégements de cotisations sociales sur l'emploi et les salaires : une évaluation de la réforme de 2003[END_REF] and Bunel and L'Horty 2019) find that the reduction in payroll taxes on the wages of the young increased growth in firms with a high share of young workers prior to the reform, relatively to firms with a lower share. 13 Symmetrically, [START_REF] Ku | Do Place-Based Tax Incentives Create Jobs?[END_REF] find that an exogenous increase in payroll taxes in some regions in Norway impulsed a significant drop in local employment, and [START_REF] Benzarti | Using Payroll Tax Variation to Unpack the Black Box of Firm-Level Production[END_REF] show that a higher payroll tax at the firm level implies (downward) employment responses, relatively to firms subject to a lower rate. More generally, my paper adds to the evidence that firms respond to positive or negative cash shocks by adjusting employment, suggesting binding financial constraints [START_REF] Melcangi | The Marginal Propensity to Hire[END_REF].

Conclusion

This paper studies a large payroll tax cut on minimum wage workers. Results show that neither changing relative labor costs for different labor types, or wages, nor the cash windfall that results from the policy, may fully explain firm behavior and the changes in the job destribution.

I find that the payroll tax cut triggered an increase in the number of jobs at the minimum wage, as was expected by the reform. However, this stems only from firms with too few low-wage workers to benefit ex ante from a reduction in labor costs thanks to the reform.

In these firms, the retention rate of low-wage workers increases, as well as the number of hires in firms with no minimum wage workers prior to the reform. Results suggest that this increase of the number of low-wage jobs is coupled with a wage trap. 13 An exception is provided by [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF] that shows that a tax cut proportional to the total payroll of employees paid less than 2.5 times the minimum wage had no effect on employment. The setting is however very different from these payroll tax cuts on two points: the tax cut is calculated based on the proportion of workers paid less than 2.5 times the minimum wage-a much higher threshold than that of the payroll tax cut studied here-and the tax cut materializes with at least a one-year delay, being a tax credit. The first feature implies that the firms benefiting from a cash windfall are very different in our settings, and in particular could differ in terms of liquidity constraints, and the demand for this higher-paid labor is suspected to be less sensitive to its cost; the second feature could further explain that firms do not take into account in year N a reduction in labor cost effective one, two or three years later.

Firms which already employed minimum-wage workers prior to the reform, and therefore benefit from an exogeneous drop in their production costs, appear not to react to the change in relative costs: on the contrary, the number of jobs at higher wages increase, through more retention of higher-wage workers and more wage increases for low-wage jobs. These firms pocket the cash windfall but do not seak to increase the benefits of this payroll tax reduction.

With a firm-level analysis, I find that these firms use the cash windfall to boost their size growth. Part of the cash windfall goes into increased profits.

From a methodological viewpoint, this paper shows the limits of focusing solely on firmlevel or job-level "treatment". Mixing both levels of analysis enables to see that both individuallevel treatment and firm scale effects matter, and moreover shows that relative-costs effects may differ according to the cash windfall firms receive.

In terms of policy, this paper shows that reducing the cost of minimum-wage labor succeeds in fostering employment at the minimum wage, but also in addressing liquidity constraints of small low-wage firms which can anticipate hiring decisions. Efficient labor cost-reducing policies should be designed in such a way that they take in account the mechanical cash windfall effect, which can help liquidity-constrained firms but can also be pocketed by lessconstrained, bigger firms, and their employees (as shown by [START_REF] Carbonnier | Who benefits from tax incentives? The heterogeneous wage incidence of a tax credit[END_REF].

Future research should investigate the conditions in which a cash windfall is used to hire and/or to increase wages of the less-skilled, low-wage workers, and those in which the cash windfall serves profits and the wages of skilled workers and managers. In particular, comparing the effects of different types of cuts in production cost, targeting different types of firms, could help to disentangle fairness motives, the importance of bargaining power of different types of workers, and the role of liquidity constraints. Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. Source: Exhaustive employer-employee data (DADS Postes). Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of continuing jobs in a given wage bin and a given year as the dependent variable. Continuing jobs are jobs that are observed both in N and in N-1. In Figure 7a, the dependent variable is the number of continuing jobs with no upward wage mobility. In Figures 7b and7c, the dependent variable is the number of continuing jobs with no upward wage mobility. In Figure 7b, the wage bin is that of the wage observed in N-1 (origin) whereas in Figure 7c, the wage bin is that of the wage observed in N (destination). Upward wage mobility is defined as a change of wage bin. Source: Exhaustive employer-employee data (DADS Postes). Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of all jobs in a given wage bin and a given year as the dependent variable. The simple differences is estimated on the sample of firms belonging to a specific quintile of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes). Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. The simple differences is estimated on the sample of firms belonging to quintile Q1 of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

- Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. The simple differences is estimated on the sample of firms belonging to quintile Q2 of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

- Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. The simple differences is estimated on the sample of firms belonging to quintile Q3 of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

- Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. The simple differences is estimated on the sample of firms belonging to quintile Q4 of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

40

- Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Net created jobs is the difference between new jobs and old jobs. New jobs are jobs that are observed in N but not in N-1; old jobs are jobs that are observed in N-1 but not in N. Continuing jobs are jobs that are observed both in N and in N-1.

The wage bin is that of the wage observed in N. The simple differences is estimated on the sample of firms belonging to quintile Q5 of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

- Note: These figures plot the β b,t estimates of the simple differences regression given by equation 1 for each bin b and each year t. The dependent variable is the number of jobs in a given wage bin b ("destination") and a given year N which were in a lower wage bin a < b in N -1. The simple differences is estimated on the sample of firms belonging to a specific quintile of exposure to the SSC reforms, with Q1 the weighted quintile of firms benefiting ex ante from a cash windfall and Q5 the weighted quintile of firms suffering the most from an ex ante increase in labor cost. Source: Exhaustive employer-employee data (DADS Postes).

- Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of the ratio of social charges in total labor costs (Figure 15a), full-year equivalent employment (Figure 15b) and profits (Figure 15c). These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the balanced panel of firms with at least 2 full-year equivalent employment in 1995 (N = 355053, 350352 and 352359, respectively). Standard errors are clustered at the firm level. The vertical bars represent the 95% confidence interval.

Source: Fiscal and employer-employee data (FICUS, DADS Postes). Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of the ratio of social charges in total labor costs (Figure A3a), full-year equivalent employment (Figure A3b) and profits (Figure A3c), with additional yearly fixed effects for quintiles of firm size in 1995. These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the balanced panel of firms with at least 2 full-year equivalent employment in 1995 (N = 355053, 350352 and 352359, respectively). Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of the ratio of social charges in total labor costs (Figure A4a), full-year equivalent employment (Figure A4b) and profits (Figure A4c). These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the balanced panel of firms with at least 5 full-year equivalent employment in 1995 (N = 460836, 446791 and 456756, respectively) Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of the ratio of social charges in total labor costs (Figure A5a), full-year equivalent employment (Figure A5b) and profits (Figure A5c). These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the balanced panel of firms with at least 10 full-year equivalent employment in 1995 (N = 96862, 96654 and 96676, respectively). Standard errors are clustered at the firm level.

The vertical bars represent the 95% confidence interval.

Source: Fiscal and employer-employee data (FICUS, DADS Postes). Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of the ratio of social charges in total labor costs (Figure A6a), full-year equivalent employment (Figure A6b) and profits (Figure A6c), with additional yearly fixed effects for quintiles of firm size in 1995. These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the balanced panel of firms with at least 10 full-year equivalent employment in 1995 (N = 96862, 96654 and 96676, respectively A7b), full-year equivalent employment (Figure A7c) and profits (Figure A7d). The regression for Figure A7b includes additional yearly fixed effects for quintiles of firm size in 1995. Full-year equivalent employment is corrected for attrition, assuming a -100% growth rate when the firm is not observed in the data. These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the unbalanced panel of firms with at least 10 full-year equivalent employment in 1995 (N = 570445, 570445, 570445 and 563059, respectively). Standard errors are clustered at the firm level. The vertical bars represent the 95% confidence interval.

Source: Fiscal and employer-employee data (FICUS, DADS Postes).

  represents the change in labor cost due to exogenous change in payroll taxes, had the labor composition of the firm remained constant between 1995 and 1997. Denoting LC the labor cost for a firm, which depends roughly on the firm and workforce characteristics W (wage distribution, executive or non-executive status, number of employees, etc.) and on the payroll tax schedule Sch, I define the ex ante change in labor cost due to the 1995-1997 reforms as: ∆LC = LC(W 1995 , Sch 1997 ) -LC(W 1995 , Sch 1995 ) LC(W 1995 , Sch 1995 ) (3) Based on this continuous measure of direct exposure to the changes in payroll taxes, I group firms into 5 or 10 groups of equal total number of workers (hereafter referred to as weighted quintiles or deciles, and denoted Q1 to Q5, D1 to D10). Firms in the first quintile or decile have the lowest, or most negative, ex ante variation of labor cost: the changes in the payroll tax schedule between 1995 and 1997 strongly decrease their labor cost. Conversely, firms in the top quintile or decile are those most affected by the change in payroll tax rates which increase labor cost ex ante.

Figure

  Figure 2a plots the distribution of ex ante change in labor costs. Labor cost can mechanically

(

  2012) highlight a strong reaction of full-time equivalent employment to cuts in labor cost in France between 2003 and 2005. Similarly, Saez et al. (

Figure 2 :

 2 Figure 2: Ex ante variation of labor cost Note: Figure 2a represents the distribution of ex ante variation of labor cost. Figures 2b and 2c plot the average values of ex ante variation of labor cost by subgroups of firms. Source: Fiscal and employer-employee data (FICUS, DADS Postes). Author's microsimulation of changes in labor costs.

Figure 3 :Figure 4 :

 34 Figure 3: Simple differences: All jobs, selected wage binsNote: This figure plots the β b,t estimates of the simple differences regression given by equation 1 for two wage bins and each year t, with a normalized measure of the number of all jobs in a given wage bin and a given year as the dependent variable. Source: Sample version of the employer-employee data (DADS Panel).

Figure 5 :Figure 6 :

 56 Figure 5: Simple difference estimates: All jobs, comparing sample and exhaustive dataNote: These figures plot a subset of the β b,t estimates of the simple differences regression given by equation 1 for bins b of width 0.2 and each year t post-1995, with a normalized measure of the number of jobs in a given wage bin and a given year as the dependent variable. Source: Sample version (panel 5a) and exhaustive (panel 5b) employer-employee data.

Figure 7 :

 7 Figure 7: Simple difference estimates: Continuing jobs with or without significant upward wage mobility

Figure 8 :

 8 Figure 8: Simple difference estimates: All jobs, by firms' quintile

Figure 9 :

 9 Figure 9: Simple difference estimates: Decomposing between job types, Q1

Figure 10 :

 10 Figure 10: Simple difference estimates: Decomposing between job types, Q2

Figure 11 :

 11 Figure 11: Simple difference estimates: Decomposing between job types, Q3

Figure 13 :

 13 Figure 13: Simple difference estimates: Decomposing between job types, Q5

Figure 14 :

 14 Figure 14: Simple difference estimates: Upwards wage mobility, destination wage bin, by firms' quintile

Figure 15 :

 15 Figure 15: Firm-level effects of changes in payroll taxes

Figure A2 :

 A2 Figure A2: Average change in employer SSCs, excluding reductions on low wages, by weighted deciles of ex ante change in labor cost

  Figure A4: Firm-level effects of changes in payroll taxes in firms with more than 5 fullyear employees in 1995

  Figure A5: Firm-level effects of changes in payroll taxes in firms with more than 10 fullyear employees in 1995

  Figure A6: Firm-level effects of changes in payroll taxes in firms with more than 10 fullyear employees in 1995: Adding firm size fixed effects

  FigureA7: Firm-level effects of changes in payroll taxes: Using an unbalanced sample Note: These figures plot the β estimates for the regression of the growth rate, since 1995, of attrition (FiguresA7a and A7b), full-year equivalent employment (FigureA7c) and profits (FigureA7d). The regression for FigureA7bincludes additional yearly fixed effects for quintiles of firm size in 1995. Full-year equivalent employment is corrected for attrition, assuming a -100% growth rate when the firm is not observed in the data. These estimates are interpreted as the differential evolution, at a particular year, of these variables for any decile with respect to decile 4. The regression is estimated on the unbalanced panel of firms with at least 10 full-year equivalent employment in 1995(N = 570445, 570445, 570445 and 563059, respectively). Standard errors are clustered at the firm level. The vertical bars represent the 95% confidence interval. Source: Fiscal and employer-employee data (FICUS, DADS Postes).

Table 1 :

 1 Number of observations Reductions in employer SSCs are deducted from the sum of all employer SSCs (pension, unemployment and non-contributory SSCs). SSC rates may differ with the employee status; the sum of employer SSCs is calculated for a non-executive employee. Source: IPP Tax and benefit tables, author's calculations.

		Nb of firms
	Data type	Balanced Unbalanced
	≥ 2 FYE in 1995	336 428	545 720
	≥ 5 FYE in 1995	176 419	264 979
	≥ 10 FYE in 1995	91 855	130 683
	FYE = workforce, full-year equivalent.

Note:

In particular, some countries such as Colombia, South Africa or Sweden have put into place payroll tax reductions for young workers. In Finland, a payroll tax subsidy exists for older workers. Reductions of labor cost are sometimes also conditional on the wage level. In Finland, firms can apply to a grant to cover parts of the labor costs when they hire a previously unemployed individual. See[START_REF] Becerra | Labor demand responses to payroll taxes in an economy with wage rigidity: Evidence from colombia[END_REF];[START_REF] Ebrahim | Can a wage subsidy system help reduce 50 per cent youth unemployment?: Evidence from South Africa[END_REF];[START_REF] Huttunen | The employment effects of low-wage subsidies[END_REF];[START_REF] Kangasharju | Do Wage Subsidies Increase Employment in Subsidized Firms[END_REF];[START_REF] Saez | Coefficient 95% Confidence Interval Figure A1: Difference in differences: a larger treatment group[END_REF].

Access to this confidential data was granted by the Comité du secret statistique, by decisions ME27 of October 2, 2013, ME56 of June 2,2014 and ME91 of June 6, 2015.

 5 To measure the size of the workforce, I also calculate the number of full-time equivalent workers in the firm using the information on the number of hours worked and contract duration of each job in the DADS. In the early years of the DADS Postes however, this measure displays poor consistency. I use the self-reported variable carried by the fiscal data instead.

The data for 1990 is no longer available, which limits the period of analysis.

Merging with the balanced sample of firms rules out the possibility for the evolution of the number of jobs to be driven by the attrition of firms observed in 1995. In particular, merging with the unbalanced firm-level data would mechanically generate an artificial spike in the number of jobs in 1995.

Note that the different effects of increases vs. decreases of labor costs will not be distinguishable from differential effects of labor cost changes on low-wage vs. high-wage workers.

Recent papers have shown that this effect of firm size on cyclical job flows is in fact entirely driven by an effect of firm age[START_REF] Haltiwanger | Who Creates Jobs? Small versus Large versus Young[END_REF][START_REF] Fort | How Firms Respond to Business Cycles: The Role of Firm Age and Firm Size[END_REF][START_REF] Colciago | Who Creates and Destroys Jobs over the Business Cycle? DNB Working Papers[END_REF].

Note: This figure plots the difference-in-differences estimates of equation 2 using a larger treatment group. Source: Sample version of the employer-employee data (DADS Panel).

I am grateful for the funding provided by France stratégie and for the helpful discussions with members of the COSAPE (Comité de suivi des aides publiques aux entreprises). Any remaining errors are my own. Data access through the Centre d'accès sécurisé aux données (CASD) has been supported by ANR grant ANR-10-EQPX-17. I acknowledge the support of the EUR grant ANR-17-EURE-0001.


Florence Biguenet
email: florence.biguenet@uha.fr

Safi Melki

Dominique Dupuis

On the Relevance to Test the Robustness of Textile Hydrophobic Properties

Controlling wetting behavior has recently attracted a great attention due to the wide applications in technologies (medical, printing, cleaning processes…). This study examines the influence of textile structures on the hydrophobicity but also on the robustness of this property. A laboratory test method to evaluate this robustness is proposed.

INTRODUCTION

Wetting properties depend on a combination of surface chemistry and surface roughness. Many studies have been performed on the effects of model textured surfaces on the hydrophobicity [START_REF] Reyssat | Impalement of fakir drops[END_REF][2] [START_REF] Marmur | Wetting on hydrophobic rough surfaces: to be heterogeneous or not to be?[END_REF]. Fewer have explored the wetting behavior of surfaces with non-perfect pattern structures as textile surfaces [START_REF] Michielsen | Design of a superhydrophobic surface using woven structures[END_REF]. Such surfaces combine different roughness scale levels related to the yarns, the fibers and sometimes to nanofibers or nanoparticles grafted on the primary structure [START_REF] Zimmermann | A Simple, One-Step Approach to Durable and Robust Superhydrophobic Textiles[END_REF]. Additionally, the different elements constituting a fabric are highly deformable.

The main characteristic of the wetting properties of non planar surfaces is the apparent contact angle. Another important characteristic is the contact angle hysteresis which is the difference between the advancing and the receding contact angle. This hysteresis is related to the pinning of the drop when the solid surface is tilted. It is proportional to the force required to start a drop moving on a surface. The present discussion focuses on the necessity to introduce another characteristic parameter to express the robustness of the hydrophobic property of rough surfaces and especially here textile surfaces. Some hydrophobic fibrous structures were prepared to prospect for the influence of textile structures on the natural and forced wetting properties.

APPROACH

All samples were submitted to the same sizing treatment in order to obtain hydrophobic surfaces. Two main classes of fabrics are considered: classical fabrics and flocked ones.

For the classical fabrics, different arrangement of the yarns and different geometries of the yarns are considered. The warp is made of cotton yarns. The weft is made of cotton, cotton/PET blend or PET yarns. PET yarns are continuous multifilament yarns while cotton and cotton/PET yarns are made of staple fibers which induces some pilosity. In order to modify the roughness at the yarn scale, two different weaves have been chosen: satin weave (SW) and plain weave (PW).

Flocked textiles are composed of a cotton substrate (canvas fabric) on which the viscose hairs of various diameters and lengths are stuck.

To measure the robustness of the hydrophobic property, a laboratory tester has been designed and produced. It allows measuring the contact angle of a water droplet before and after compression on the surface.

RESULTS AND DISCUSSION

The classical fabrics are characterized by several scales of roughness corresponding respectively to the crosslinks of the yarns (weaving), the yarn structure and the fibers geometry. Our woven samples, exhibiting approximately the same natural contact angle (125° to 130°), show significant different forced wetting behavior. The fibers geometry and the weaving structure seems to complementary influence the contact angle hysteresis and the robustness of the hydrophobic property. The sateen weave clearly displays the influence of the type of weft yarn on the hysteresis and the robustness of the hydrophobic property. Indeed, the contact area of the liquid drop with these yarns is more important than in the case of plain weave.

The pile fabrics are characterized by a greater contact angle (142° to 147°) and ability for the drop to roll on it.

Nevertheless, when an external pressure is applied to the drop, the pile samples lose their hydrophobicity with contact angles being able to decrease up to 50°. The drop is, at rest, in a Cassie state and undergoes a wetting transition to the Wenzel state after compression. But the problem is more complicated than for model surfaces since the hairs are deformable and bend during the compression which significantly modifies the initial structure. As observed for the woven samples, the pile samples, of approximatively the same static wetting properties, appear to resist differently to compression.

CONCLUSION

The measure of the static wetting properties (natural contact angle, hysteresis, roll-off angle) is not sufficient to characterize the wetting dynamic behavior of a textile surface. Such measurements have to be completed by a study on the influence of an external stress. Here, a laboratory test method is proposed to evaluate the robustness of hydrophobicity. This property can be characterized by the smaller pressure required to induce the lost of hydrophobicity.

KEYWORDS

Contact Angle, Hydrophobicity, Contact Angle Hysteresis, Forced Wetting, Wetting Transition.

Fig. 1 :

 1 Fig. 1: Example of a drop before and after compression


Bernd Wittmann

Stephan Wiesneth

Sajedeh Motamen

Laurent Simon

Françoise Serein-Spirau

Günter Reiter

Richard Hildner

Energy transport and light propagation mechanisms in organic single crystals

niveau recherche, publiés ou non, émanant des établissements d'enseignement et de recherche français ou étrangers, des laboratoires publics ou privés.