Willingness to Pay for a Saltwater Recreational Fishing License: A Comparison of Angler Groups

Transcription

1 Marine Resource Economics, Volume 16, pp /00 $ Printed in the U.S.A. All rights reserved Copyright 2002 Marine Resources Foundation Willingness to Pay for a Saltwater Recreational Fishing License: A Comparison of Angler Groups JOHN C. WHITEHEAD University of North Carolina at Wilmington WILLIAM B. CLIFFORD THOMAS J. HOBAN North Carolina State University Abstract We consider the proposed saltwater recreational fishing license in North Carolina and compare three samples of recreational anglers: fishing club members, commercial license holders without an endorsement to sell, and intercepted anglers. Anglers are faced with a dichotomous choice willingness-to-pay question with multiple follow-ups by telephone and in-person surveys. While support for the saltwater license as a management tool is relatively low, most anglers are willing to purchase a license if it is required for fishing and if the funds are used to improve fishing quality. Fishing club members have the highest willingness to pay among the three groups. Several differences in the determinants of willingness to pay emerge. An estimate of the potential revenue from the fishing license fee is $18 million. An estimate of the potential net benefit from improved fishing quality resulting from management activities funded by the license fee is $71 million. Key words Dichotomous choice with follow-ups, random effects probit model, saltwater recreational fishing license, willingness to pay. Introduction North Carolina s coastal fisheries management system has not adequately determined and responded to the needs of its anglers, fish stock, and fisheries habitats partly because of rapid demographic change and population growth. The 1997 North Carolina Fisheries Reform Act (FRA) addressed these needs. The FRA developed a new licensing structure, reorganized the North Carolina Marine Fisheries Commission, required development of Fishery Management and Coastal Habitat Protection Plans, and increased penalties for fisheries violations. A saltwater recreational fishing license, the controversial Coastal Recreational Fishing License (CRFL), was an John C. Whitehead is an associate professor of economics in the Department of Economics and Finance, University of North Carolina at Wilmington, 601 South College Road, Wilmington, NC , whiteheadj@uncwil.edu. William B. Clifford is a professor in the Department of Sociology and Anthropology, North Carolina State University, Box 8107, Raleigh, NC , william_clifford@ncsu.edu. Thomas J. Hoban is a professor in the Department of Sociology and Anthropology, North Carolina State University, Box 8107, Raleigh, NC , tom_hoban@ncsu.edu. The North Carolina Sea Grant Program funded this study (Project No. 98-FEG-40). Contents of the publication do not necessarily reflect the views and policies of the North Carolina Sea Grant Program, and the mention of any tradenames or commercial products does not constitute their endorsement by the North Carolina government. We thank Ron McPherson, Jack Thigpen, and Vernon Kelley for their contributions to this research. A previous version of this paper was presented at the 14 th Annual Meeting of the Tidewater Chapter of the American Fisheries Society in Kill Devil Hills, North Carolina. 177

2 178 Whitehead, Clifford, Hoban important piece of the proposed new license structure. However, it was not part of the enacted legislation. Debate over the CRFL occurred without the data necessary to fully understand the economic issues involved (Griffith 1996). During 1998, we conducted the Coastal North Carolina Fishing Survey (Clifford et al. 1999). The purpose of the survey was to better understand the knowledge and attitudes of coastal recreational anglers about a range of issues affecting the fishery resources. At the time of the survey, the North Carolina General Assembly was considering an annual saltwater recreational fishing license with a cost of $15 and a one-week license with a cost of $5. Revenue generated by the CRFL would be used by the North Carolina Division of Marine Fisheries (DMF) to improve the quality of saltwater recreational fishing. In the survey, we paid particular attention to the economic evaluation of this version of the CRFL. The research design employed was cross-sectional utilizing random sampling procedures. Since random samples of specialized populations, such as saltwater recreational anglers, are difficult to obtain without license lists (Teisl and Boyle 1997), we contacted a broad group of individuals involved in saltwater recreational fishing: (i) saltwater fishing club members, (ii) commercial fishing license holders without endorsements to sell, and (iii) saltwater anglers along the coast of North Carolina during the fall of None of the samples are representative of the population of saltwater recreational anglers in North Carolina. We estimate willingness to pay (WTP) for the CRFL with the stipulation that license revenues are devoted to fisheries management activities designed to improve fishing quality. We use dichotomous choice contingent valuation with multiple follow-up questions. The contingent valuation method is an approach to estimating behavioral intentions and economic value under hypothetical situations (Mitchell and Carson 1989). It is especially useful when the behavior of interest is beyond the range of historical experience (e.g., purchase of a saltwater recreational fishing license in North Carolina). In the typical dichotomous choice approach, survey respondents are presented with a hypothetical program or policy description and a price, randomly varied across respondents, and then asked whether they would purchase or vote for the price/policy combination. The variation in the price allows estimation of maximum WTP for the policy. Our application does not vary the price across respondents, since it was well known at the time of the survey that the proposed annual license fee was $15. Respondents who state that they would not purchase the annual license were asked if they would be willing to pay $5 for the one-week CRFL. One of our primary goals was to determine how many anglers would purchase the annual and one-week CRFLs at the expected fees. In order to estimate the WTP for quality improvements associated with the CRFL, respondents who stated that they would purchase the CRFL for $15 were presented with a follow-up valuation question with a higher fee, randomly varied across respondents. Additional follow-up valuation questions were asked until WTP was bounded by yes and no responses or an upper bound was reached. The maximum WTP can be estimated using the initial and follow-up questions using panel data and the typical dichotomous choice empirical approach (Cameron and James 1987). Previous research has found that there is a tendency for contingent valuation respondents to answer no to dichotomous choice follow-up questions due to incentive incompatibility (Alberini, Kanninen, and Carson 1997). In the present context, the follow-up questions might suggest to respondents that increases in the license fee will lead to government waste or that government may not be able to provide the quality improvement at lower fees. Risk averse respondents will be more likely to answer no. Similar to past research (Alberini, Kanninen, and Carson 1997), we test for an incentive incompatibility effect on the follow-up responses. In contrast,

3 WTP for a Recreational Fishing License 179 we compare a uniform incentive incompatibility effect with a decomposition of the effect into its explanatory variables. In the next section, we describe the survey methodology and the three samples of recreational anglers. Next, we compare the demographics and fishing characteristics of these anglers. Then we present the WTP and related variables and a comparison of these data across angler groups. We describe the theoretical model of WTP and its estimation with the random effects probit model for panel data. A comparison of angler groups by regression model and WTP is then presented. Conclusions, policy implications, and suggestions for future research follow. The Data Three populations of anglers were sampled: (i) saltwater fishing club members, (ii) commercial fishing license holders without endorsements to sell, and (iii) saltwater anglers along the coast of North Carolina during the fall of The twelve-month project began in August 1998 precluding on-site interviews with anglers during the spring and summer fishing seasons. Based on the lack of a representative sample frame, we suspected that our sample would be more avid than a representative sample of North Carolina anglers. A systematic random sample of fishing club members was drawn from the membership lists of the following clubs in proportion to their representation of the total club members: Piedmont Offshore Sport Fishing Club, Oak Island Fishing Club, Raleigh Saltwater Sport Fishing Club, Charlotte Sport Fishing Club, Winston-Salem Saltwater Sport Fishing Club, and Core Banks Surf Fishing Club. This sample was surveyed by telephone. A systematic random sampling procedure was also used to generate a sample from commercial fishing license holders who did not possess an endorsement to sell seafood in 1998 (referred hereafter as license holders ). In North Carolina, this group of recreational anglers is allowed to use commercial gear (e.g., gill nets, trawls, or crab pots) but cannot sell their catch. 1 The population list (n = 3,700) was obtained from the North Carolina DMF. This sample was also surveyed by telephone. A stratified sampling scheme was developed to select intercept interview sites based on fishing mode and region. Six fishing modes were selected to encompass the diversity of saltwater recreational fishing in North Carolina: (i) public and wildlife accesses, (ii) beach/surf, (iii) marinas, (iv) piers, (v) banks and bridges, and (vi) charter and head boats. Fifty interviews from each mode were conducted. A further refinement in the sampling scheme employed regional sampling based on dividing the North Carolina coast into northern, central, and southern regions. Each of the six modes was represented in each region. Specific sites for the identified fishing modes were determined using available lists, including: 1997 North Carolina Boating Guide, Marine Recreational Fisheries Statistics Survey Master Site Register, and North Carolina topographical maps. Convenience sampling was conducted at each site. Sampling was conducted mostly on weekends. 2 1 In 1999 the license system was changed as required by the 1997 FRA. The license for the commercial fishing license holders is now called the Recreational Commercial Gear License. It is an annual license that allows recreational fishermen to use limited amounts of commercial gear to harvest seafood for their personal consumption. Seafood harvested under this license cannot be sold. Fishermen using this license will be held to recreational size and possession limits. recgear.htm. Authorized gear includes gill nets, pots, shrimp trawls, trotlines, and seines. The cost of the annual license is $35. 2 Additional details regarding site selection and sampling for the intercept sample population are provided in Clifford et al. (1999).

4 180 Whitehead, Clifford, Hoban Telephone surveys were completed with 121 fishing club members and 203 license holders during October to December of An overall response rate of 84% was obtained for the two samples by dividing the number of completed interviews by the total number contacted (completed, terminated, and refused). A total of 300 in-person interviews were completed with anglers during October, November, and December of A response rate of 77% was obtained for the in-person sample by dividing the number completed by the total number contacted (completed, terminated, and refused). 3 Fishing characteristic variables suggest that the respondents in all three samples are avid anglers (table 1). The average, annual number of trips taken in North Carolina is greater than 24 for each group. Each group of anglers has been fishing in saltwater for greater than one-half of their lives. The average years of fishing experience is greater than 23 for each group. These statistics support our suspicion that our respondents are more avid than the general population of saltwater anglers in North Carolina. The populations from which the samples are drawn are not mutually exclusive. Five percent and 14% of the anglers in the license holder and intercept samples belong to fishing clubs. Nineteen percent and 4% of the anglers in the fishing club and intercept samples have a commercial fishing license. However, none of the anglers in the fishing club or intercept samples report using commercial gear, while 78% of the license holder sample reports using commercial gear. Ninety-eight percent, 59%, and 94% of the fishing club, commercial license, and intercept sample anglers, respectively, report using hook and line, the most common recreational gear. A small portion of the sample (6.5%) is composed of out-of-state anglers. Twelve percent of the intercept sample is from out-of-state, while only 3% and 1% of the fishing club and license holder samples are from out-of-state. We do not discard these respondents from the analysis since the out-of-state sub-sample is large enough (n = 40) to provide useful information about the WTP of out-of-state anglers. A statistical comparison using differences in means tests of the demographic variables indicates that the three groups of anglers are distinctly different (table 1). Respondents in the license holder sample have more fishing experience than fishing club anglers (t = 3.70) and the intercept anglers (t = 6.44). Those in the fishing club sample have higher family incomes than respondents in the license holder (t = 7.47) and intercept (t = 10.28) samples. Commercial fishing license holders are older than those in the fishing club sample (t = 4.26) and the intercept sample (t = 9.24). Those in the fishing club sample have obtained more education than those in the license holder (t = 7.98) and intercept (t = 6.63) samples. Finally, using the Pearson chisquare statistic, those in the intercept sample are more likely to be non-white and female relative to those in the fishing club (χ 2 = 6.58[1 df], χ 2 = 9.10[1 df]) and license holder (χ 2 = 6.58 [1 df], χ 2 = 9.10 [1 df]) samples. Respondents were also asked whether they support or oppose a series of management alternatives related to the FRA, including a saltwater recreational fishing license. More than 75% of each sample support regulations such as creel and size limits. More than two-thirds of each sample support land protection policies, such as purchasing buffer strips along waterways and stronger wetlands protection laws. The majority of respondents in each sample support quotas and seasons. However, less 3 The in-person response rate is an upper bound. We found it difficult to obtain interviews from the target number of charter boat anglers (n = 25). The interviewers failed to keep an adequate count of the number of charter boat anglers who refused the interview. This does not greatly affect the response rate for this sample. Assuming a small undercount of charter boat interview refusals (e.g., 2 boats, n = 12) leads to a response rate of 75%.

5 WTP for a Recreational Fishing License 181 Table 1 Demographic Variables Fishing Club License Holders Intercept Name Description Mean STD Mean STD Mean STD TRIPS Trips during past 12 months YEARSFSH Years fishing in saltwater FISHCLUB = 1 if belongs to fishing club, 0 otherwise COMMLCNS = 1 if possesses commercial license, 0 otherwise INCOME 1998 family income AGE 1998 minus birth year EDUC Highest grade of school completed WHITE = 1 if race is white, 0 otherwise MALE = 1 if gender is male, 0 otherwise

6 182 Whitehead, Clifford, Hoban than 50% of the license holder and intercept angler samples support a saltwater recreational fishing license as a management tool. Eighty percent of the fishing club sample supports the license. Using the Pearson chi-square statistic, the differences between the fishing club and license holder (χ 2 = 39.37[1 df]) and intercept (χ 2 = 47.56[1 df]) samples are statistically significant. Coastal Recreational Fishing License In the contingent valuation section of the survey, respondents were told that the CRFL would be required for anyone who fishes in North Carolina coastal waters and that all the money raised would go into a dedicated fund (see Appendix). Money from the fund would be used to improve saltwater recreational fishing through habitat restoration, better enforcement, expanded research, and education. Respondents were also told that the license would help to obtain an accurate count of the number of recreational anglers. Then, anglers were asked if they would purchase the annual CRFL at the $15 fee. Anglers who stated that they would not purchase or did not know whether they would purchase the annual CRFL were asked if they would purchase the one-week CRFL at a $5 fee. In contrast to the lack of support for the license as a management tool, a significant majority of anglers are willing to pay for the CRFL if it is required to fish and is associated with a quality improvement (table 2). Ninety-one percent, 68%, and 79% of the fishing club member, license holder, and intercept angler samples, respectively, would purchase the annual CRFL. These differences are statistically significant according to the Pearson chi-square statistic (χ 2 = 23.69[2 df]). Zero, 14%, and 37% of the remaining fishing club, license holder, and intercept anglers, respectively, would purchase the one-week CRFL. These differences are statistically significant according to the Pearson chi-square statistic (χ 2 = 6.09[2 df]). Altogether, 84% of the anglers would purchase either the one-week or annual CRFL. The anglers (n = 101) who would not purchase either the annual or one-week CRFL were then asked to name all of the reasons why. Almost one-half (n = 46) stated that they, don t support a saltwater fishing license. The next most common reason was that, the license fee is not fair (n = 42). Other popular reasons were that they, don t trust government (n = 21), they already pay for a freshwater license (n = 18), commercial fishers should pay (n = 13), management won t in- Table 2 First and Second Willingness-to-Pay Responses by Fee Amount Fishing Club License Holder Intercept Yes (%) Total Yes (%) Total Yes (%) Total A 1 $ $ $ $ $ $ $

7 WTP for a Recreational Fishing License 183 crease fish (n = 13), and commercial boats catch all the fish (n = 11). Many of these responses suggest that these anglers are protesting the notion of a saltwater recreational fishing license or rejecting the contingent valuation scenario. We include these respondents in the analysis recognizing that the effect is to potentially bias the willingness-to-pay estimates downward. Anglers who state that they would purchase the annual CRFL at $15 were then told to suppose that it would take more money to improve fisheries quality and asked whether they would be willing to purchase the annual CRFL at a higher cost. Each angler was presented a different license fee: $25, $35, $50, $75, or $ We find significant differences in WTP higher fees across angler groups (table 2). According to the Cochran-Mantel-Haenszel chi-square statistic, the fishing club members are most willing to pay higher fees, holding constant the fee amount (χ 2 = [1 df]). Using the Pearson chi-square statistic, the differences in the probability of a yes response are statistically significantly across fee amounts for the fishing club (χ 2 = 11.70[4 df]), license holder (χ 2 = [4 df]), and intercept (χ 2 = 53.77[4 df]) sample anglers. More than 50% of the fishing club sample answer yes to each additional follow-up question (table 3). Fifty percent or less of the other samples answer yes to each additional follow-up question. Holding the panel number constant, the fishing club members are significantly more likely to respond yes to the follow-up questions when compared to the license holder (χ 2 = 28.88[1 df]) and intercept (χ 2 = 44.39[1 df]) angler samples using the Cochran-Mantel-Haenszel chi-squared statistic. There is no difference in the proportion of follow-up yes responses between the license holder and intercept samples (χ 2 = 0.38[1 df]). In order to determine the reasons why respondents are willing to pay for the CRFL, they were asked whether they strongly agree, agree, disagree, or strongly disagree with statements about the license. The statements measure respondent attitudes about whether: the money raised would help fisheries, the license is a good idea if funds are used in the sounds or estuaries, the license would improve saltwater fishing in the same way freshwater fishing had improved, the license would provide better information for fisheries managers, recreational fishermen need to help pay to manage the fishery, and it is important to keep track of the number of recreational anglers. We create dummy variables that are equal to one if the respondent either agreed or strongly agreed with the statement and zero otherwise. Seventy-nine percent and greater of the fishing club sample strongly agrees or agrees with each of the statements (table 4). Between 55% and 78% or less of license holder and intercept samples strongly agree or agree with each of the statements. Pearson chi-square statistics for no association between the dummy variable and angler sample were computed for each variable. When comparing the fishing club sample to each of the other samples separately, all of the differences in proportions are statistically significant (p = 0.05). Few differences emerge between the license holder and intercept sample anglers. A significantly greater number of intercept anglers think that money raised by the license would help the fisheries in North Carolina (p = 0.05) and a license would provide better information (p = 0.10). 4 The bid amounts were selected considering the tradeoff between realism and coverage of the range of willingness to pay. Since it was widely known that $15 was the proposed cost of the annual fishing license, higher amounts would suffer from a lack of realism. Our initial list of bid amounts included two amounts greater than $100 ($150 and $200). The advisory committee on the project, including anglers and social scientists, recommended dropping these amounts, which we did. In the telephone survey, the Computer Assisted Telephone Interview software chose the bids randomly. In the on-site interview, each interviewer chose the bids sequentially with a random start at each site. According to our survey design, the average fee amounts do not differ across sample.

8 184 Whitehead, Clifford, Hoban Table 3 Willingness-to-Pay Responses by Panel Fishing Club License Holder Intercept Panel Yes (%) Total Yes (%) Total Yes (%) Total Table 4 Attitudes about a Saltwater Recreational Fishing License Percent Who Agree or Disagree Fishing License Variable Statement Club Holder Intercept HELP FUNDS FRESH INFO NEED TRACK Money raised by a saltwater fishing license would help the fisheries in North Carolina A saltwater fishing license is a good idea if the funds were used in North Carolina s sounds or estuaries Because money raised by the freshwater license has helped improve recreational fishing, it will do the same for saltwater fishing A saltwater fishing license would provide better information for fisheries managers Recreational fishermen need to pay to help manage the fishery It is important to keep track of how many people fish in saltwater Willingness to Pay Model Anglers possess the utility function u(z, x, q) which is increasing in z, the composite of all market goods; increasing in x, saltwater recreational fishing trips; and increasing in q, fishing quality. The expenditure function, e(p, c, F, q, u), results from the expenditure minimization problem: {min [pz + cx + F] s.t. u = u(z, x, q)} where p is the composite price of all market goods, c is the per trip cost of fishing, and F is the license fee. The expenditure function is increasing in p, c, F, and u and decreasing in q. Purchasing the CRFL is required for saltwater recreational fishing. Holding quality constant, the WTP for access to the recreational fishery is: WTP e p c Fc C, = (,,, q, u) e( p, c, F = 0, q, u). (1) F F Equation (1) defines WTP as the difference in two expenditure functions. The first expenditure function measures the expenditures required to reach the reference level of utility when the annual fee is the choke price, F c, and, assuming perfect enforcement, no recreational fishing trips are taken. The second is the expenditures required to reach the reference level of utility under the status quo.

9 WTP for a Recreational Fishing License 185 Revenue from the CRFL would be used to improve fishing quality. The WTP for the quality improvement is: WTPqq, * = e( p, c, F = 0, q, u) e( p, c, F, q*, u). (2) The first expenditure function represents the status quo. The second expenditure function measures expenditures when q* is improved quality and F > 0. Note that the WTP for the quality improvement could be less than or equal to zero. The WTP for both characteristics of the CRFL, fishing access with improved quality, is the summation of equations (1) and (2): c WTPCRFL = WTPF c, F + WTPqq, * = e( p, c, F, q, u) e( p, c, F, q*, u) (3) where the status quo expenditure functions cancel. In equation (3), anglers compare the expenditures necessary to reach the reference level of utility without fishing access and the expenditures necessary to reach the reference level of utility with a positive annual fee and improved quality. WTP is greater than or equal to zero since F c > F and q < q*. The maximum WTP for the CRFL is estimated using data from the dichotomous choice with follow-up contingent valuation questions. The choice for the respondent is to first decide whether the annual value of saltwater recreational fishing trips with improved quality is greater than the CRFL fee, $A 1 = $15: yes 1 = 1 if WTPCRFL $ 15 0 if WTPCRFL < $ 15 (4) If the respondent answers no, he is presented with the one-week license offered for $A 2 = $5. The follow-up decision for no respondents is: yes 2 = 1 if WTPCRFL $ 5 0 if WTPCRFL < $ 5 (5) If the respondent answers yes to the first question, he is presented with a hypothetical situation where improved quality requires higher fee revenue. The respondent is then asked to pay a higher fee, $A f, in a series of follow-up questions, f = 2,, 6. The follow-up decisions are: yes f = 1 0 if WTP $ A CRFL if WTP < $ A CRFL f f (6) The possible values for $A f are $25, $35, $50, $75, and $100. If the respondent answers yes to a follow-up question, the next question is repeated with the next highest fee until the respondent answers no or the $100 fee is reached. If the respondent answers no to a follow-up question, the next question is repeated with the next lowest value until the respondent answers yes or the $25 fee is reached. We treat the dichotomous choice with follow-up data as a panel and estimate the probability of a yes response with the random effects probit model:

10 186 Whitehead, Clifford, Hoban Pr( yes ) = Pr( WTP A ) = Pr( β X + e A ) (7) = it it t i it t Pr β Xi At eit β Xi At = Φ σ σ σ υ υ υ where X i is a vector of variables, β is a vector of corresponding coefficients, i = 1,, n respondents, and t = 1,, f i observations (questions) per respondent. The error term, e it = u i + υ it, is the sum of the error term common to the individual, u i, and the random error term, υ it. The correlation coefficient between the responses, ρ = σ u2 /(σ u 2 + σ u2 ), is a measure of the ratio of the variance in the random effects component of the model to the total variance in the model. The larger the correlation, the more likely the random effects model is appropriate (Greene 2000, pp ). To test for incentive incompatibility, we include a variable equal to one for the follow-up responses in panels 2 through 6. 5 We estimate the following specification: Pr( yes ) it = + XXi + DD At Φ β β β 0 σ υ (8) where β = [β 0, β X, β D ] and D is a dummy variable equal to 1 if t > 1 and 0 if t = 1. A negative and statistically significant coefficient on D indicates incentive incompatibility in the follow-up questions. Alternatively, we interact the dummy variable for the follow-up questions with the vector of independent variables: Pr( yes ) it = + XXi + DD + XD( Xi D) At Φ β β β β 0. σ υ (9) A comparison of the two models reveals whether the incentive incompatibility effect is uniform or heterogeneous across respondents. When D = 0, this model simulates the responses with incentive compatibility: + XXi At Pr( yesi ) = Φ β β 0. σ υ (10) The coefficients on the WTP model are estimated from the censored probit coefficients using the procedures described in Cameron and James (1987). Since the dollar amount is varied across respondents, σ υ can be identified and WTP can be recovered: WTP CRFL = β0 + β X X (11) The standard errors for WTP CRFL are constructed using the Delta Method (Cameron 1991; Greene 2000, p. 278). 5 In initial models, separate variables were included for the follow-up responses for those who answered yes and no to the initial question at the $15 bid for each group. Likelihood ratio test statistics indicate that the coefficients on these variables are not significantly different from each other. Therefore, we collapse these variables into a single variable for a follow-up response. We also test for anchoring on the randomly assigned bid amounts given to respondents who answer yes to the first valuation question. No evidence of anchoring is found.

11 WTP for a Recreational Fishing License 187 Empirical Results We first estimate pooled random effects probit models (table 5). The measure of the contribution of the individual error terms to the total error term (ρ) is almost 60% in each model, indicating that the random effects specification is appropriate. Differences in the three samples are examined with dummy variables for the whether the respondent belongs to a fishing club and possesses a commercial fishing license. Note that these dummy variables, which allow testing for the effects of club membership and owning a commercial license in the split-sample models, are different from fishing club and license holder sample dummy variables due to the non-exclusivity of the samples. Models that included dummy variables for the fishing club and license holder samples do not yield qualitatively different results in the pooled model. The license fee, the dummy variable for follow-up questions, years fishing Table 5 Random Effects Probit Models: Pooled Sample Model 1 Model 2 Model 3 Variable Coeff. t-ratio Coeff. t-ratio Coeff. t-ratio Constant FEE (A) FOLLOWUP (D) YEARSFSH YEARSFSH INCOME HELP FUNDS FRESH INFO NEED TRACK FISHCLUB COMMLCNS D*YEARSFSH D*YEARSFSH D*INCOME D*HELP D*FUNDS D*FRESH D*INFO D*NEED D*TRACK D*FISHCLUB D*COMMLCNS ρ Beginning LL 1, , , Ending LL 1, , , χ 2 : Model 1 vs. Model 2 [11 df] χ 2 : Model 1 vs. Model 3 [5 df] Observations 2,167 2,167 2,167 Cases

12 188 Whitehead, Clifford, Hoban and its square, income, and attitudes about the saltwater fishing license are included as explanatory variables in each of the models. 6 A primary result is that the coefficient on the license fee variable is negative and statistically significant (Model 1). Anglers are less likely to state that they would purchase the CRFL as the fee increases. The coefficient on the dummy variable for the follow-up questions is negative and statistically significant, indicating that the follow-up questions are not incentive compatible. Anglers are more likely to purchase the CRFL with increases (at a decreasing rate) in fishing experience. Anglers with higher incomes are more likely to purchase the CRFL, indicating that it is a normal good. Most of the coefficients on the attitudinal variables are positive and statistically significant. Anglers who think that the license revenues would improve fishing, the license is a good idea if the fees are dedicated to improving fishing quality, and recreational fishermen need to contribute to fisheries management are more likely to state that they would purchase the CRFL. Anglers who belong to a fishing club are more likely to be willing to pay for the license. Anglers who own a commercial fishing license are less likely to be willing to pay for the recreational license. In Model 2, we include a vector of interaction variables equal to the dummy for the follow-up questions multiplied by all independent variables other than the license fee. These interaction variables decompose the follow-up question effect and help identify those anglers for which the follow-up questions are incentive incompatible. Most of the results on the independent variables are the same as in Model 2, except the coefficient on the dummy variable for the follow-up question is no longer statistically significant. Also, the coefficients on the dummy variable for fishing club membership and the attitudinal variable about the license improving the fishery are no longer statistically significant. Finally, the standard errors of several variables in Model 2 are larger than in Model 1. According to the likelihood ratio statistic, the vector of interaction coefficients is statistically significant at the p = 0.05 level, indicating that the size of the incentive incompatibility effect varies across anglers (χ 2 = 23.48[11 df]). Anglers with more fishing experience (at a decreasing rate) are less likely to answer yes in the follow-up questions. Two attitudinal variables related to the license revenues have interaction coefficients that are negative and statistically significant. Respondents who agree with the statements about how the funds should be used and the similarity between the effectiveness of the saltwater and freshwater licenses are less likely to answer yes in the follow-up questions. In contrast, respondents who agree that it is important to keep track of the number of recreational anglers are more likely to answer yes in the follow-up questions. Commercial license holders are also more likely to answer yes in the follow-up questions. In Model 3, we drop the interaction variables that do not have statistically significant coefficients. The results indicate that Model 3 is the preferred model. According to the likelihood ratio statistic, the vector of interaction coefficients is statistically significant at the p = 0.01 level relative to Model 1. Another likelihood ratio statistic indicates that Model 3 is preferred over Model 2 (χ 2 = 2.58[5 df]). The standard errors of those variables that increased in Model 2 are reduced in Model 3. The coefficients on the variables that became statistically insignificant in Model 2 are no longer statistically insignificant. The split-sample versions of Model 3 are presented in table 6. The result of the likelihood-ratio test for the appropriateness of the split-samples (χ 2 = 56.95[20 df]) 6 We do not include the variables measuring age, education, gender, race, and out-of-state anglers in the empirical models. Of these, only the race variable is statistically significant (with whites willing to pay more) in the pooled models. Since the sample of non-white anglers in the fishing club sample is too small to obtain a reliable coefficient estimate, we do not include this variable in any of the models.

13 WTP for a Recreational Fishing License 189 indicates that the coefficients should not be constrained equal in the pooled Model 3. The measure of the contribution of the individual error terms to the total error term (ρ) ranges from 59% to 65% in the models, indicating that the random effects specification is appropriate. In each model, the coefficient on the license fee variable is negative and statistically significant. In the fishing club sample model, only three other variables have statistically significant coefficients. Anglers with higher incomes are more likely to respond yes. Anglers are more likely to respond yes if they equate the potential results of the saltwater license with those of the freshwater license and if they think recreational anglers need to help pay for fisheries management. None of the coefficients on the follow-up variables are statistically significant. The likelihood ratio statistic indicates that the vector of follow-up variables is not statistically significant (χ 2 = 10.59[11 df]). However, when all of the interaction variables are dropped from the model, the coefficient on the follow-up dummy variable is negative and statistically significant at the p = 0.01 level. This result indicates that the size of the incentive incompatibility effect is uniform across anglers in the fishing club sample. In the license holder sample, anglers who think that license revenues will improve the quality of fishing, equate the potential results of the saltwater license with those of the freshwater license, and think recreational anglers need to help pay for fisheries management are more likely to be willing to pay for the CRFL. However, anglers who equate the potential results of the saltwater license with those of the freshwater license are less likely to answer yes in the follow-up questions. Anglers who think it is important to keep track of the number of recreational anglers are more likely to respond yes to the follow-up questions. Table 6 Random Effects Probit Models: Split-Sample Fishing Club License Holder Intercept Variable Coeff. t-ratio Coeff. t-ratio Coeff. t-ratio Constant FEE (A) FOLLOWUP (D) YEARSFSH YEARSFSH INCOME HELP FUNDS FRESH INFO NEED TRACK FISHCLUB COMMLCNS D*YEARSFSH D*YEARSFSH D*FUNDS D*FRESH D*TRACK D*COMMLCNS ρ Beginning LL Ending LL Observations ,061 Cases

14 190 Whitehead, Clifford, Hoban The intercept angler sample model closely resembles the pooled Model 3 in table 5. Intercept anglers are more likely to purchase the CRFL with increases (at a decreasing rate) in fishing experience. Those with higher incomes and those who are members of a fishing club are more likely to purchase the CRFL. Intercept anglers who own a commercial fishing license are less likely to be willing to pay for the recreational license. Anglers who think that the license revenues would improve fishing, the CRFL is a good idea if the license fees are dedicated to improving fishing quality, and recreational fishermen need to contribute to fisheries management are more likely to be willing to pay. Anglers with more fishing experience are less likely to answer yes to the follow-up questions (at a decreasing rate). Respondents who agree with the statements about how the funds should be used and the similarity between the effectiveness of the saltwater and freshwater licenses are less likely to answer yes in the follow-up questions. Anglers who agree that it is important to keep track of the number of recreational anglers are more likely to answer yes in the follow-up questions. Intercept sample anglers who are commercial license holders are more likely to answer yes in the follow-up questions. Willingness to pay is calculated by setting the dummy variable for the follow-up question equal to zero, simulating incentive compatible follow-up questions (table 7). The WTP estimates from the three split-sample models suggest that the differences between the maximum WTP for the CRFL and the proposed cost are large for each sub-sample. There are some differences in the WTP from the three groups. The fishing club sample WTP point estimate is about 200% larger than those of the license holder and intercept sample anglers. However, the lack of precision in the fishing club WTP estimate leads to a wide 90% confidence interval. This points to a conclusion of no statistically significant difference between the WTP of the fishing club and license holder samples at the p = 0.10 level. The fishing club and intercept angler samples WTP estimates are significantly different at the p = 0.10 level. In order to obtain WTP estimates with tighter confidence intervals, we next estimate simple random effects probit models with only the license fee and dummy variable for follow-up questions as independent variables. 7 In each case, the WTP estimate from the simple model is lower than the WTP estimate from the full model (table 7). However, none of the differences between the full and simple models are statistically significant. The differences in WTP between the fishing club sample and the license holder and intercept samples are statistically significant at the p = 0.10 level. Conclusions In this paper, we estimate the WTP for access to improved saltwater recreational fishing in North Carolina with a license fee payment vehicle and dichotomous choice with follow-up valuation questions. While there is relatively low support for a saltwater recreational fishing license in two of our three samples of anglers, there is a clear willingness to purchase the $15 annual license when anglers are told that the money generated from license sales will be used to improve the quality of the fishery. We find that 84% of all anglers would purchase either the annual or one-week license. Furthermore, we find positive attitudes about the impact of a license. There is sentiment that the li- 7 These results are available upon request.

15 WTP for a Recreational Fishing License 191 Table 7 Willingness-to-Pay Estimates Full Model Fishing Club License Holder Intercept Mean $ $62.63 $66.51 (2.92) a (4.30) (7.26) 90% C.I. Lower Bound $81.96 $38.61 $ % C.I. Upper Bound $ $86.65 $81.63 Simple Model Mean $ $52.09 $60.23 (4.13) (5.16) (8.29) 90% C.I. Lower Bound $81.37 $35.42 $ % C.I. Upper Bound $ $68.76 $72.23 a t-ratio in parentheses. cense would help the fisheries in North Carolina, that recreational anglers need to pay to help manage the fishery, and that the license would provide better information for fisheries management. Using panel data models, we find that the determinants of the responses to the WTP questions differ significantly across the fishing club, license holder, and intercept angler samples. The overwhelming willingness of fishing club sample anglers to purchase the license leads to few statistically significant determinants of WTP. License holders and intercept sample anglers are more likely to purchase the CRFL if they have positive attitudes about the fishery. Intercept sample anglers with higher incomes and more fishing experience are more likely to purchase the CRFL. Two results are common to all angler groups. All anglers are more likely to be willing to pay if they think that the saltwater license would have benefits similar to the freshwater license, and recreational anglers should help fund fisheries management. Consistent with previous research, we find that the follow-up questions are not incentive compatible. In contrast to previous research, we find that the total effect can be parameterized into its components for two of the three samples. Anglers who agree that the license is a good idea if the money is used for management in the sounds and estuaries are less likely to respond yes in followup questions. One possible explanation of this result is as follows. Since the focus of the use of the revenues is limited to the sounds and estuaries, excluding ocean enforcement activities for example, anglers may consider raising extra revenue with a higher license fee wasteful. Anglers who agree that the saltwater license will have benefits similar to those of the freshwater license are also less likely to respond yes in follow-up questions. Since the benefits from the freshwater license are generated from a low fee, anglers may be suspicious of a higher saltwater license fee. Anglers who agree that it is important to track the number of recreational anglers are more willing to pay in the follow-up questions. Perhaps anglers feel this is a justifiable reason for raising extra revenue. While our samples are not representative of the population of North Carolina saltwater anglers, we can use these results to get a general idea of the outcomes of a saltwater recreational fishing license. The revenue from the CRFL would be used to finance fishery management programs. An estimate of the amount of money potentially available for recreational fisheries management is

16 192 Whitehead, Clifford, Hoban the $15 fee multiplied by the number of anglers who purchase the annual license plus the $5 fee multiplied by the number of one-week anglers who purchase the one-week license. Using angler estimates from the period (North Carolina DMF 1997) and linear extrapolation, we project the number of in-state and out-ofstate anglers for the years The projection of the number of in-state and out-of-state anglers in 2000 is 616,308 and 887,342 (Clifford et al. 1999). Seventy-nine percent of the most representative of our samples, the intercept angler sample, would purchase the annual CRFL, and 7.4% would purchase the one-week CRFL. If 79% of all anglers purchase the annual license and 7.4% purchased the one-week license, $18 million in new revenues would be available to fund fisheries management programs. The WTP estimates indicate that the benefits of a saltwater fishing license are large for each angler group. Using the more conservative estimates from the simple models, WTP for the annual license is $135, $52, and $60 for the fishing club, license holder, and intercept sample anglers, respectively. The annual net benefit of the CRFL is equal to WTP minus the cost of the license. Using the WTP estimate from the intercept sample, the percentage of intercept anglers who would purchase the annual CRFL, and the aggregate angler estimates from above, an estimate of the annual net benefit of improved fishing trips due to the CRFL is $71 million. Several caveats should be kept in mind. First, we suspect the intercept sample is subject to avidity bias. We expect the revenue and net benefit estimates to be biased upward as a result. Second, it is questionable whether a number of the respondents who answer no to both valuation questions are giving truthful responses. Their reasons for answering these questions suggest that they reject the contingent valuation scenario or protest the policy. These anglers state that they would stop fishing rather than purchase the license. If these anglers would actually purchase the CRFL if placed in that situation, leaving these respondents in the sample biases the revenue and WTP estimates downward. Finally, the proportion of in-state and out-of-state anglers in our sample is not representative of the true proportion. We have only 11% out-ofstate anglers in the intercept sample when the true proportion is about 60%. While we find no difference in the proportion of yes responses among in-state and out-of-state anglers, a more representative sample might find otherwise. During the 1999 session of the North Carolina General Assembly, the missing piece of the proposed North Carolina FRA was again considered. Debate centered on whether the magnitude of the license fee would discourage fishing participation. In response, the proposed annual fee was reduced to $7.50. During the 2000 and 2001 legislative sessions, proponents of the CRFL again pursued its passage and argued that there is wide support among recreational anglers for a license. Their efforts have failed largely due to political pressure from the commercial fishing industry, which fears the potential revenue generated by the CRFL and the accompanying political power of recreational anglers. In the absence of passage of a saltwater recreational fishing license in North Carolina, future research should continue to explore the WTP for the CRFL. Our research is limited, since time and budget constraints led to avid samples from unrepresentative angler populations. Future research should be conducted with a more representative angler sample. For example, the sample should include anglers who fish during the spring and summer seasons as well as the fall season. Also, by sampling during the summer tourist season, a larger proportion of out-of-state anglers would be interviewed.