Abstract
Even if there exists an extensive literature on the modeling of farmers’ behavior under risk, actual measurements of the quantitative impact of risk aversion on input use are rare. In this article, we use simulations to quantify the impact of risk aversion on the optimal quantity of input and farmers’ welfare when production risk depends on how much of the input is used. The assumptions made on the technology and form of farmers’ risk preferences were chosen such that they are fairly representative of crop farming conditions in the USA and Western Europe. In our benchmark scenario featuring a traditional expected utility model, we find that less than 4% of the optimal pesticide expenditure is driven by risk aversion and that risk induces a decrease in welfare that varies from −1.5 to −3.0% for individuals with moderate to normal risk aversion. We find a stronger impact of risk aversion on quantities of input used when farmers’ risk preferences are modeled under the cumulative prospect theory framework. When the reference point is set at the median or maximum profit, and for some levels of the parameters that describe behavior toward losses, the quantity of input used that is driven by risk preferences represents up to 19% of the pesticide expenditure.
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Notes
Some empirical evidence of pesticides reducing production risk were found in Di Falco and Chavas [22] using data from durum wheat farms in Sicily (Italy) and in Koundouri et al. [32] on a panel of Finnish crop farms but Serra et al. [57] and Serra et al. [58], in a study of 596 farms from Kansas, found pesticides to be risk-increasing inputs. This discrepancy in findings may be explained by the effect of inputs varying depending on whether crop growth conditions are good or bad [59] and/or by the use of different types of indicators of pesticide application [8].
The author does not indicate how the cost of risk compares to the expected profit and certainty equivalent.
In these articles, technology parameters and the parameters representing farmers’ risk preferences (often a risk aversion coefficient) were estimated within a unique model made of simultaneous equations, in most cases under the assumption that farmers face production risk only. Separate identification of the technology and risk preferences in such models was later called into question by Lence [60] and Just and Just [61].
In line with the assumption of narrow bracketing [62], we assume pesticide decisions are made in isolation from other decisions made on the farm. Among the determinants of narrow bracketing are two factors that are relevant for pesticide use: farmers face cognitive resource scarcity and use heuristics (treatment planning) that increase the probability of narrow bracketing in pesticide use. We are aware of only one study pointing to narrow bracketing by farmers in their choice of inputs. Using an investment game to experimentally elicit risk preferences, Verschoor et al. [63] show that risk preferences in the lab significantly explain fertilizer use of Ugandan crop farmers, but not the decision to grow cash crops or the degree of market participation. They suggest that this result can be explained by narrow bracketing in fertilizer use, a decision made in isolation and similar to the decision in the experimental investment game.
For calibration of the production function and in order to determine plausible ranges for pesticide use and prices, we relied on observational data from farms producing cereals in northeastern France. The major crops produced on these farms were wheat, barley, and rapeseed. We used annual data on winter wheat over the period 1993–2010 (see Appendix A1). For additional information on these data, see also Carpentier and Letort [64], Femenia and Letort [45], and Koutchadé et al. [65].
Expenditure is often the only information that is available to researchers, in particular for those using data from the (European) Farm Accountancy Data Network. However, and as will be discussed in Sect. 5.3., more refined indicators of pesticide use are recommended when available [8].
Yields can only be positive or null so the generated yield is set at zero when realizations of the random shock produce a negative yield.
Empirical estimates of the relative risk aversion coefficient vary on a wide range and estimated coefficients around 6 are not uncommon (e.g., [30, 31, 38], cf. Appendix A3).
The empirical analysis has been made using R. Programs are available in the supplementary material.
Risk-loving farmers decrease their use of pesticides if pesticides are risk decreasing. When the relative risk aversion coefficient is set at -1, the optimal input use is 5 EUR lower than in the risk-neutral case and the risk premium is negative in this case and equal to 16 EUR. With a relative risk aversion coefficient of -2, the negative risk premium reaches 31 EUR (there was a positive risk premium of 34 EUR under risk aversion with a relative risk aversion coefficient of + 2). We observe the same phenomenon (impact of same magnitude and opposite sign) also for relative risk aversion coefficients equal to -3 and -4.
Historical data for the price of wheat (in EUR) are publicly available here: https://www.indexmundi.com/commodities/?commodity=wheat&months=180¤cy=eur
These variations in pesticide use in response to price changes would correspond to price elasticities of demand for pesticide varying between −0.7 and −1.2, which is in the lower end of the distribution of elasticity estimates measured in the literature [66].
Using data from French cereal crop farmers, Femenia and Letort [45] found that a 35% pesticide tax would induce a 25% reduction in pesticide use. The tax rate that is necessary to induce a 25% reduction in pesticide use is slightly higher in their case but we believe it is partly driven, among other differences in the modelling framework, by their assumption of a higher output price. In their simulations they set the price of wheat at 171 EUR per ton (against 110 EUR per ton in our case).
When rare catastrophic weather events occur and damage the crops very badly, farmers usually benefit from specific disaster funds that help compensate for losses. Hence farmers in Europe do not really have any incentive to manage the risk of (catastrophic) crop damage and losses.
There is rare theoretical guidance for the choice of the reference point and the need to make assumptions on where such a reference point is seen as one of the major weaknesses of CPT Barberis [44].
We also ran one scenario featuring Tversky and Kahneman (12)’s parameters except for loss aversion being set at 3.50. The optimal input use with the reference point being set at 0, the median profit, and the maximum profit is 270 EUR, 279 EUR, and 279 EUR, respectively. This scenario is not shown in Tables 4 and 5 due to space constraints.
A few countries have adopted pesticide taxation, the most commonly cited being Sweden and Denmark. Sweden applies a tax to each kilogram of active substance (EUR 3.64/kg in 2016) while Denmark applies a pesticide tax that is differentiated by the pesticides’ category (insecticides, herbicides, fungicides, and growth regulators). For greater information on pesticide taxation schemes, we refer readers to Böcker and Finger (67) and Finger et al. (68).
When rare catastrophic weather events occur and damage the crops very badly, farmers usually benefit from specific disaster funds that help compensate for losses. Hence farmers in Europe do not really have any incentive to manage the risk of (catastrophic) crop damage and losses.
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Funding
The study is financially supported by the project “Facilitate public Action to exit from peSTicides (FAST)” as part of the French Priority Research Programme “Growing and Protecting Crops Differently of the French National Research Agency (ANR) and the French National Research Agency (ANR) under the Investments for the Future (Investissements d’Avenir) program, grant ANR-17-EURE-0010.
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Appendices
Appendix 1
1.1 Description of the Crop Data from France Used for Calibration of the Production Function
We use data from crop farms from the Meuse département (a small administrative region in northeastern France). Our sample is a rotating panel which covers years from 1993 to 2010. The major crops produced on these farms are wheat, barley, and rapeseed. In the following, we focus on winter wheat, which was grown on 1,144 farms on our sample over the entire period (with the number of observations per year varying from 255 to 687 farms). We focus on a single-output single-input function, and we choose pesticides as the major input of the production of winter wheat. Information on input use is crop-specific and is measured in real monetary terms (expenditure). Below are two graphs featuring smoothed regressions of wheat yield on pesticides expenditure:
As expected, the relationship is increasing and concave. We try and fit a Just-Pope function on these data. The best fit was obtained with the following function: \(y = 16 \times {x^{0.28}} + 30 \times {x^{ - 0.20}} \times \varepsilon\), which we approximate with \(y = 15 \times {x^{0.30}} + 30 \times {x^{ - 0.20}} \times \varepsilon\) in our simulation exercise. The price of wheat is set at 110€/t, which corresponds to the median on our sample over the entire period.
Appendix 2
2.1 Review of Technology Parameters: Elasticities of Mean Output and Output Variance to Input Use
We report below estimates from the literature of the elasticity of mean output and output variance to input use, for studies that have estimated production functions and measured output in physical quantities. All these estimates are for arable farms producing crops.
Article | Country and inputs | Elasticity of mean output | Elasticity of output variance |
---|---|---|---|
Just and Pope [21] | Country: USA Input: fertilizers | 0.3‐0.4 | 0.1‐0.2 |
Gardebroek et al. [46] | Country: Netherlands Input: herbicides and pesticides | 0.23 (for conventional farms) | 0.21 (but not statistically significant) |
Koundouri et al. [32] | Country: Finland Inputs: fertilizers and plant protection | 0.50 for plant protection 0.77 for fertilizers | −0.05 for plant protection −0.02 for fertilizers |
Femenia and Letort [45] | Country: France Inputs: fertilizers and pesticides | 0.1 for both fertilizers and pesticides | - |
Appendix 3
3.1 Review of Relative Risk Aversion Parameters Derived from Expected Utility Models
Authors | Sample | Coefficient of relative risk aversion |
---|---|---|
Bontems and Thomas [24] | Corn plots from US Midwest 1990–1992 | 3.7 |
Chavas and Holt [31] | US aggregate corn and soybean acreage decisions 1954–1985 | [1.4–6.8] |
Isik and Khanna [47] | Field-level data (99 fields) from Illinois (USA) 1993–1994 | 1.5 |
Lence [48] | US aggregate data 1934–1994 | [1.1–2.5] |
Love and Buccola [29] | Farm data on corn and soybean production, Iowa (USA) | [2.4–18.8] |
Pope et al. [49] | State-level data on net returns and acreages, 8 US states | 0.4 |
Saha et al. [30] | Farm-level data for wheat farms in Kansas (USA), 1979–1982 | [3.8–5.4] |
Sckokai and Moro [38] | Farm-level data from Italy, 1993–1999 | [0–5.5] |
Appendix 4
4.1 Sensitivity of the Optimal Input Use to the Mean Function
Appendix 5
5.1 Sensitivity of Input Use and Risk Premium to Levels of Relative Risk Aversion and Parameters of the Risk Function
Risk neutrality | Risk aversion | |||||
---|---|---|---|---|---|---|
Risk function | rr = 0 | rr = 1 | rr = 2 | rr = 3 | rr = 4 | |
Risk-decreasing input | ||||||
\(g(x) = 20{x^{ - 0.1}}\) | Input use (EUR) | 264 | 266 | 268 | 270 | 272 |
Self-insurance (%) | 0.0 | 0.8 | 1.5 | 2.2 | 2.9 | |
RP* (EUR) | 0 | 7 | 15 | 22 | 30 | |
\(\Delta\) CE (%) | - | −0.6 | −1.3 | −2.0 | −2.6 | |
\(g(x) = 40{x^{ - 0.1}}\) | Input use (EUR) | 264 | 273 | 283 | 295 | 309 |
Self-insurance (%) | 0.0 | 3.3 | 6.7 | 10.5 | 14.6 | |
RP* (EUR) | 0 | 30 | 63 | 100 | 142 | |
\(\Delta\) CE (%) | - | −2.7 | −5.7 | −9.0 | −12.7 | |
\(g(x) = 30{x^{ - 0.1}}\) | Input use (EUR) | 264 | 268 | 273 | 279 | 285 |
(Benchmark) | Self-insurance (%) | 0.0 | 1.5 | 3.3 | 5.4 | 7.4 |
RP* (EUR) | 0 | 17 | 34 | 52 | 71 | |
\(\Delta\) CE (%) | - | −1.5 | −3.0 | −4.6 | −6.3 | |
\(g(x) = 30{x^{ - 0.2}}\) | Input use (EUR) | 264 | 267 | 270 | 273 | 276 |
Self-insurance (%) | 0.0 | 1.1 | 2.2 | 3.3 | 4.3 | |
RP* (EUR) | 0 | 5 | 11 | 16 | 21 | |
\(\Delta\) CE (%) | - | −0.5 | −1.0 | −1.4 | −1.9 | |
Risk-increasing input | ||||||
\(g(x) = 10{x^{0.1}}\) | Input use (EUR) | 264 | 259 | 253 | 247 | 241 |
Self-insurance (%) | 0.0 | −1.9 | −4.3 | −6.9 | −9.5 | |
RP* (EUR) | 0 | 17 | 35 | 54 | 73 | |
\(\Delta\) CE (%) | - | −1.5 | −3.1 | −4.8 | −6.6 | |
\(g(x) = 15{x^{0.1}}\) | Input use (EUR) | 264 | 251 | 234 | 213 | 187 |
Self-insurance (%) | 0.0 | −5.2 | −12.8 | −23.9 | −41.2 | |
RP* (EUR) | 0 | 41 | 87 | 142 | 207 | |
\(\Delta\) CE (%) | - | −3.6 | −7.8 | −12.8 | −18.7 |
Appendix 6
6.1 Sensitivity to Initial Wealth
Risk neutrality | Risk aversion | |||||
---|---|---|---|---|---|---|
Initial wealth | rr = 0 | rr = 1 | rr = 2 | rr = 3 | rr = 4 | |
500 EUR | Input use (EUR) | 264 | 268 | 273 | 279 | 285 |
(benchmark) | Self-insurance (%) | 0.0 | 1.5 | 3.3 | 5.4 | 7.4 |
RP* (EUR) | 0 | 17 | 34 | 52 | 71 | |
\(\Delta\) CE (%) | - | −1.5 | −3.0 | −4.6 | −6.3 | |
1000 EUR | Input use (EUR) | 264 | 267 | 270 | 273 | 277 |
Self-insurance (%) | 0.0 | 1.1 | 2.2 | 3.3 | 4.7 | |
RP* (EUR) | 0 | 11 | 23 | 34 | 46 | |
\(\Delta\) CE (%) | - | −0.7 | −1.4 | −2.1 | −2.8 | |
1500 EUR | Input use (EUR) | 264 | 266 | 268 | 271 | 273 |
Self-insurance (%) | 0.0 | 0.8 | 1.5 | 2.6 | 3.3 | |
RP* (EUR) | 0 | 8 | 17 | 26 | 34 | |
\(\Delta\) CE (%) | - | −0.4 | −0.8 | −1.2 | −1.6 | |
5000 EUR | Input use (EUR) | 264 | 264 | 265 | 266 | 267 |
Self-insurance (%) | 0.0 | 0 | 0.4 | 0.8 | 1.1 | |
RP* (EUR) | 0 | 3 | 6 | 10 | 13 | |
\(\Delta\) CE (%) | − | −0.1 | −0.1 | −0.2 | −0.2 |
Appendix 7
7.1 Review of Parameters Derived from CPT Models
Authors, sample | Specification | Parameter estimates |
---|---|---|
Tversky and Kahneman [12] 25 Students | CRRA utility function No preference reversal \({\alpha }^{+}={\alpha }^{-}=\alpha\) Loss aversion \(\lambda\) KT probability weighting function \({\gamma }^{+}\ne {\gamma }^{-}\) | \(\widehat{\alpha }=\widehat{{\alpha }^{+}}=\widehat{{\alpha }^{-}}=0.88\) (Farmers are risk averse) \(\widehat{\lambda }\approx 2.25\) (Farmers are loss averse) \(\widehat{{\gamma }^{+}}\approx 0.61;\widehat{{\gamma }^{-}}\approx 0.69\) (Less probability distortion in losses than in gains) |
Tanaka et al. [50] 181 Villagers, Vietnam | CRRA utility function No preference reversal \({\alpha }^{+}={\alpha }^{-}=\alpha\) Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=0.60\) \(\widehat{\lambda }=3.47\) \(\widehat{\gamma }=0.74\) |
Nguyen and Leung [51] 103 Fishermen, Vietnam | CRRA utility function No preference reversal α+ = α− Loss aversion λ Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=0.62\) \(\widehat{\lambda }=2.63\) \(\widehat{\gamma }=0.75\) |
Harrison et al. [52] 531 Villagers, Ethiopia, India and Uganda | CRRA utility function No loss aversion (gains only) KT probability weighting function γ | \(\widehat{\alpha }=0.54\) \(\widehat{\gamma }=1.38\) |
CRRA utility function No loss aversion (gains only in the experiment) Prelec probability weighting function \(\gamma\) (and additional scale parameter \(\eta\)) | \(\widehat{\alpha }=0.50\) \(\widehat{\gamma }=0.96\) \(\widehat{\eta }=1.20\) | |
Nguyen [53] 181 Fishermen, Vietnam | CRRA utility function No preference reversal α+ = α− = α Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=1.01\) \(\widehat{\lambda }=3.26\) \(\widehat{\gamma }=0.96\) |
Liu and Huang [54] 320 Cotton farmers, China | CRRA utility function No preference reversal α+ = α− = α Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=0.52\) \(\widehat{\lambda }=3.47\) \(\widehat{\gamma }=0.69\) |
Bocquého et al. [14] Farmers, France | CRRA utility function No preference reversal α+ = α− = α Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=0.28\) \(\widehat{\lambda }=2.28\) \(\widehat{\gamma }=0.66\) |
Liebenehm and Waibel [55] 211 Cattle farmers, Mali and Burkina Faso | CRRA utility function No preference reversal α+ = α− = α Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}={\gamma }^{-}=\gamma\) | \(\widehat{\alpha }=0.11\) \(\widehat{\lambda }=1.35\) \(\widehat{\gamma }=0.13\) |
Bougherara et al. [15] 197 Crop farmers, France | CRRA utility function No preference reversal α+ = α− = α Loss aversion \(\lambda\) KT probability weighting function \({\gamma }^{+}\ne {\gamma }^{-}\) | \(\widehat{\alpha }=0.61\) \(\widehat{\lambda }=1.37\) \(\widehat{{\gamma }^{+}}=0.79;{\gamma }^{-}=0.84\) |
CRRA utility function No preference reversal \({\alpha }^{+}={\alpha }^{-}=\alpha\) Loss aversion \(\lambda\) Prelec probability weighting function \({\gamma }^{+}\ne {\gamma }^{-}\) | \(\widehat{\alpha }=0.63\) \(\widehat{\lambda }=1.38\) \(\widehat{{\gamma }^{+}}=0.81;\widehat{{\gamma }^{-}}=0.89\) | |
CRRA utility function Preference reversal \({\alpha }^{+}\ne {\alpha }^{-}\) No loss aversion KT probability weighting function \({\gamma }^{+}\ne {\gamma }^{-}\) | \(\widehat{{\alpha }^{+}}=0.60;\widehat{{\alpha }^{-}}=0.66\) \(\widehat{{\gamma }^{+}}=0.79;\widehat{{\gamma }^{-}}=0.84\) |
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Bontemps, C., Bougherara, D. & Nauges, C. Do Risk Preferences Really Matter? The Case of Pesticide Use in Agriculture. Environ Model Assess 26, 609–630 (2021). https://doi.org/10.1007/s10666-021-09756-8
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DOI: https://doi.org/10.1007/s10666-021-09756-8