1. Microeconometric Modeling
William Greene
Stern School of Business
New York University
New York NY USA
2.1 Binary Choice Models

2. Concepts Models Random Utility Maximum Likelihood Parametric Model
Partial Effect
Average Partial Effect
Odds Ratio
Linear Probability Model
Cluster Correction
Pseudo R squared
Likelihood Ratio, Wald, LM
Decomposition of Effect
Exclusion Restrictions
Incoherent Model
Nonparametric Regression
Klein and Spady Model
Probit
Logit
Bivariate Probit
Recursive Bivariate Probit
Multivariate Probit
Sample Selection
Panel Probit

3. Central Proposition A Utility Based Approach
Observed outcomes partially reveal underlying preferences
There exists an underlying preference scale defined over alternatives, U*(choices)
Revelation of preferences between two choices labeled 0 and 1 reveals the ranking of the underlying utility
U*(choice 1) > U*(choice 0) Choose 1
U*(choice 1) < U*(choice 0) Choose 0
Net utility = U = U*(choice 1) - U*(choice 0). U > 0 => choice 1

4. Binary Outcome: Visit Doctor In the 1984 year of the GSOEP, 1611 of individuals visited the doctor at least once.

5. A Random Utility Model for the Binary Choice
Yes or No decision | Visit or not visit the doctor
Model: Net utility of visit at least once
Net utility depends on observables and unobservables
Udoctor = Net utility = U*visit – U*not visit
Udoctor =  + 1Age + 2Income + 3Sex + 
Choose to visit at least once if net utility is positive
Observed Data: X = Age, Income, Sex
y = 1 if choose visit,  Udoctor > 0, 0 if not.
Random Utility

6. Modeling the Binary Choice Between the Two Alternatives
Net Utility Udoctor = U*visit – U*not visit
Udoctor =  + 1 Age + 2 Income + 3 Sex + 
Chooses to visit: Udoctor > 0
 + 1 Age + 2 Income + 3 Sex +  > 0
Choosing to visit is a random outcome because of 
 > -( + 1 Age + 2 Income + 3 Sex)

7. Probability Model for Choice Between Two Alternatives People with the same (Age,Income,Sex) will make different choices between  is random. We can model the probability that the random event “visits the doctor”will occur.
Probability is governed by , the random part of the utility function.
Event DOCTOR=1 occurs if  > -( + 1Age + 2Income + 3Sex)
We model the probability of this event.

8. An Application 27,326 Observations in GSOEP Sample 1 to 7 years, panel
7,293 households observed
We use the 1994 year; 3,337 household observations

9. Binary Choice Data (all years)

10. An Econometric Model Choose to visit iff Udoctor > 0
Udoctor =  + 1 Age + 2 Income + 3 Sex + 
Udoctor > 0   > -( + 1 Age + 2 Income + 3 Sex)  <  + 1 Age + 2 Income + 3 Sex)
Probability model: For any person observed by the analyst,
Prob(doctor=1) = Prob( <  + 1 Age + 2 Income + 3 Sex)
Note the relationship between the unobserved  and the observed outcome DOCTOR.

11. Index = +1Age + 2 Income + 3 Sex Probability = a function of the Index. P(Doctor = 1) = f(Index)
Internally consistent probabilities: (1) (Coherence) < Probability < 1 (2) (Monotonicity) Probability increases with Index.

12. Econometric Issues
Data may reveal information about coefficients I.e., effect of observed variables on utilities
May reveal information about probabilities I.e., probabilities under certain assumptions
Data on choices made do not reveal information about utility itself
The data contain no information about the scale of utilities or utility differences.
Variance of  is not estimable so it is normalized at 1 or some other fixed (known) constant.

13. Modeling Approaches Nonparametric Regressions
P(Doctor=1)=f(Income)
P(Doctor=1)=f(Age)
Essentially, at each value of Income or Age, examine the proportion of individuals who visit the doctor.

14. Klein and Spady Semiparametric Model No specific distribution assumed
Prob(yi = 1 | xi ) = G(’x) G is estimated by kernel methods
Note necessary normalizations. Coefficients are relative to FEMALE.

15. A Fully Parametric Model
Index Function: U = β’x + ε
Observation Mechanism: y = 1[U > 0]
Distribution: ε ~ f(ε); Normal, Logistic, …
Maximum Likelihood Estimation: Max(β) logL = Σi log Prob(Yi = yi|xi)
We will focus on parametric models
We examine the linear probability “model” in passing.

16. A Parametric Logit Model
We examine the model components.

17. Parametric Model Estimation
How to estimate , 1, 2, 3?
The technique of maximum likelihood
Prob[doctor=1] = Prob[ > -( + 1 Age + 2 Income + 3 Sex)]
Prob[doctor=0] = 1 – Prob[doctor=1]
Requires a model for the probability

18. Completing the Model: F()
The distribution
Normal: PROBIT, natural for behavior
Logistic: LOGIT, allows “thicker tails”
Gompertz: EXTREME VALUE, asymmetric
Others…
Does it matter?
Yes, large difference in estimates
Not much, quantities of interest are more stable.

19. Estimated Binary Choice Models for Three Distributions
Log-L(0) = log likelihood for a model that has only a constant term.
Ignore the t ratios for now.

20. Effect on Predicted Probability of an Increase in Age
 + 1 (Age+1) + 2 (Income) + 3 Sex (1 is positive)

21. Partial Effects in Probability Models
Prob[Outcome] = some F(+1Income…)
“Partial effect” = F(+1Income…) / ”x” (derivative)
Partial effects are derivatives
Result varies with model
Logit: F(+1Income…) /x = Prob * (1-Prob)  
Probit:  F(+1Income…)/x = Normal density  
Extreme Value:  F(+1Income…)/x = Prob * (-log Prob)  
Scaling usually erases model differences

23. Estimated Partial Effects for Three Models (Standard errors to be considered later)

24. Partial Effect for a Dummy Variable Computed Using Means of Other Variables
Prob[yi = 1|xi,di] = F(’xi+di) where d is a dummy variable such as Sex in our doctor model.
For the probit model, Prob[yi = 1|xi,di] = (x+d),  = the normal CDF.
Partial effect of d
Prob[yi = 1|xi, di=1] Prob[yi = 1|xi, di=0] =

25. Partial Effect – Dummy Variable

26. Computing Partial Effects
Compute at the data means (PEA)
Simple
Inference is well defined.
Not realistic for some variables, such as Sex
Average the individual effects (APE)
More appropriate
Asymptotic standard errors are slightly more complicated.

27. Partial Effects

28. Average Partial Effects Partial Effects at Data Means
The two approaches often give similar answers, though sometimes the results differ substantially.
Average Partial Effects Partial Effects at Data Means

29. APE vs. Partial Effects at the Mean

33. Odds Ratios
This calculation is not meaningful if the model is not a binary logit model

34. Odds Ratio
Exp() = multiplicative change in the odds ratio when z changes by 1 unit.
dOR(x,z)/dx = OR(x,z)*, not exp()
The “odds ratio” is not a partial effect – it is not a derivative.
It is only meaningful when the odds ratio is itself of interest and the change of the variable by a whole unit is meaningful.
“Odds ratios” might be interesting for dummy variables

35. The Linear Probability “Model”

36. The Dependent Variable equals zero for 98. 9% of the observations
The Dependent Variable equals zero for 98.9% of the observations. In the sample of 163,474 observations, the LHS variable equals 1 about 1,500 times.

37. 2SLS for a binary dependent variable.

38. Average Partial Effects
OLS approximates the partial effects, “directly,” without bothering with coefficients.
MLE
Average Partial Effects
OLS Coefficients

39. Modeling a Binary Outcome
Did firm i produce a product or process innovation in year t ? yit : 1=Yes/0=No
Observed N=1270 firms for T=5 years,
Observed covariates: xit = Industry, competitive pressures, size, productivity, etc.
How to model?
Binary outcome
Correlation across time
Heterogeneity across firms

40. Application

41. Probit and LPM

42. How Well Does the Model Fit the Data?
There is no R squared for a probability model.
Least squares for linear models is computed to maximize R2
There are no residuals or sums of squares in a binary choice model
The model is not computed to optimize the fit of the model to the data
How can we measure the “fit” of the model to the data?
“Fit measures” computed from the log likelihood
Pseudo R squared = 1 – logL/logL0
Also called the “likelihood ratio index”
Direct assessment of the effectiveness of the model at predicting the outcome

43. Pseudo R2 = Likelihood Ratio Index

44. The Likelihood Ratio Index
Bounded by 0 and a number < 1
Rises when the model is expanded
Specific values between 0 and 1 have no meaning
Can be strikingly low even in a great model
Should not be used to compare models
Use logL
Use information criteria to compare nonnested models
Can be negative if the model is not a discrete choice model. For linear regression, logL=-N/2(1+log2π+log(e’e/N)]; Positive if e’e/N < 44

45. Fit Measures Based on LogL
Binary Logit Model for Binary Choice
Dependent variable DOCTOR
Log likelihood function Full model LogL
Restricted log likelihood Constant term only LogL0
Chi squared [ 5 d.f.]
Significance level
McFadden Pseudo R-squared – LogL/logL0
Estimation based on N = , K = 6
Information Criteria: Normalization=1/N
Normalized Unnormalized
AIC LogL + 2K
Fin.Smpl.AIC LogL + 2K + 2K(K+1)/(N-K-1)
Bayes IC LogL + KlnN
Hannan Quinn LogL + 2Kln(lnN)
Variable| Coefficient Standard Error b/St.Er. P[|Z|>z] Mean of X
|Characteristics in numerator of Prob[Y = 1]
Constant| ***
AGE| ***
AGESQ| ***
INCOME|
AGE_INC|
FEMALE| ***

46. Fit Measures Based on Predictions
Computation
Use the model to compute predicted probabilities
P = F(a + b1Age + b2Income + b3Female+…)
Use a rule to compute predicted y = 0 or 1
Predict y=1 if P is “large” enough
Generally use 0.5 for “large” (more likely than not)
Fit measure compares predictions to actuals
Count successes and failures

47. Cramer Fit Measure +----------------------------------------+
| Fit Measures Based on Model Predictions|
| Efron = |
| Ben Akiva and Lerman = |
| Veall and Zimmerman = |
| Cramer = |

48. Hypothesis Tests We consider “nested” models and parametric tests
Test statistics based on the usual 3 strategies
Wald statistics: Use the unrestricted model
Likelihood ratio statistics: Based on comparing the two models
Lagrange multiplier: Based on the restricted model.
Test statistics require the log likelihood and/or the first and second derivatives of logL

49. Computing test statistics requires the log likelihood and/or standard errors based on the Hessian of LogL

50. Robust Covariance Matrix (Robust to the model specification
Robust Covariance Matrix (Robust to the model specification? Latent heterogeneity? Correlation across observations? Not always clear)

51. Robust Covariance Matrix for Logit Model Doesn’t change much
Robust Covariance Matrix for Logit Model Doesn’t change much. The model is well specified.
| Standard Prob % Confidence
DOCTOR| Coefficient Error z |z|>Z* Interval
Conventional Standard Errors
Constant| ***
AGE| ***
AGE^2.0| ***
INCOME|
|Interaction AGE*INCOME
_ntrct02|
FEMALE| ***
Robust Standard Errors
Constant| ***
AGE| ***
AGE^2.0| ***
INCOME|
_ntrct02|
FEMALE| ***

52. Base Model for Hypothesis Tests
Binary Logit Model for Binary Choice
Dependent variable DOCTOR
Log likelihood function
Restricted log likelihood
Chi squared [ 5 d.f.]
Significance level
McFadden Pseudo R-squared
Estimation based on N = , K = 6
Information Criteria: Normalization=1/N
Normalized Unnormalized
AIC
Variable| Coefficient Standard Error b/St.Er. P[|Z|>z] Mean of X
|Characteristics in numerator of Prob[Y = 1]
Constant| ***
AGE| ***
AGESQ| ***
INCOME|
AGE_INC|
FEMALE| ***
H0: Age is not a significant determinant of Prob(Doctor = 1)
H0: β2 = β3 = β5 = 0

53. Likelihood Ratio Test Null hypothesis restricts the parameter vector
Alternative relaxes the restriction
Test statistic: Chi-squared = 2 (LogL|Unrestricted model – LogL|Restrictions) > 0
Degrees of freedom = number of restrictions

54. Chi squared[3] = 2[-2085.92452 - (-2124.06568)] = 77.46456
LR Test of H0: β2 = β3 = β5 = 0
UNRESTRICTED MODEL
Binary Logit Model for Binary Choice
Dependent variable DOCTOR
Log likelihood function
Restricted log likelihood
Chi squared [ 5 d.f.]
Significance level
McFadden Pseudo R-squared
Estimation based on N = , K = 6
Information Criteria: Normalization=1/N
Normalized Unnormalized
AIC
RESTRICTED MODEL
Binary Logit Model for Binary Choice
Dependent variable DOCTOR
Log likelihood function
Restricted log likelihood
Chi squared [ 2 d.f.]
Significance level
McFadden Pseudo R-squared
Estimation based on N = , K = 3
Information Criteria: Normalization=1/N
Normalized Unnormalized
AIC
Chi squared[3] = 2[ ( )] =

55. Wald Test of H0: β2 = β3 = β5 = 0
Unrestricted parameter vector is estimated
Discrepancy: q= Rb – m is computed (or r(b,m) if nonlinear)
Variance of discrepancy is estimated: Var[q] = R V R’
Wald Statistic is q’[Var(q)]-1q = q’[RVR’]-1q

56. Wald Test
Chi squared[3] =

57. Multivariate Binary Choice Models
Bivariate Probit Models
Analysis of bivariate choices
Marginal effects
Prediction
No bivariate logit – there is no reasonable bivariate counterpart
Simultaneous Equations and Recursive Models
A Sample Selection Bivariate Probit Model
The Multivariate Probit Model
Specification
Simulation based estimation
Inference
Partial effects and analysis
The ‘panel probit model’

58. Application: Health Care Usage
German Health Care Usage Data, 7,293 Individuals, Varying Numbers of Periods Variables in the file are Data downloaded from Journal of Applied Econometrics Archive. This is an unbalanced panel with 7,293 individuals. They can be used for regression, count models, binary choice, ordered choice, and bivariate binary choice.  This is a large data set.  There are altogether 27,326 observations.  The number of observations ranges from 1 to 7.  (Frequencies are: 1=1525, 2=1079, 3=825, 4=926, 5=1051, 6=1000, 7=887).  Note, the variable NUMOBS below tells how many observations there are for each person.  This variable is repeated in each row of the data for the person. 
DOCTOR = 1(Number of doctor visits > 0) HOSPITAL = 1(Number of hospital visits > 0)
HSAT =  health satisfaction, coded 0 (low) - 10 (high)  
DOCVIS =  number of doctor visits in last three months HOSPVIS =  number of hospital visits in last calendar year PUBLIC =  insured in public health insurance = 1; otherwise = ADDON =  insured by add-on insurance = 1; otherswise = 0
HHNINC =  household nominal monthly net income in German marks /
(4 observations with income=0 were dropped) HHKIDS = children under age 16 in the household = 1; otherwise = EDUC =  years of schooling
AGE = age in years
MARRIED = marital status
EDUC = years of education

59. The Bivariate Probit Model

60. ML Estimation of the Bivariate Probit Model

61. Application to Health Care Data
x1=one,age,female,educ,married,working
x2=one,age,female,hhninc,hhkids
BivariateProbit ; lhs=doctor,hospital
; rh1=x1
; rh2=x2;marginal effects $

62. Parameter Estimates
FIML Estimates of Bivariate Probit Model
Dependent variable DOCTOR HOSPITAL
Log likelihood function
Estimation based on N = , K = 12
Variable| Coefficient Standard Error b/St.Er. P[|Z|>z] Mean of X
|Index equation for DOCTOR
Constant| ***
AGE| ***
FEMALE| ***
EDUC| ***
MARRIED|
WORKING| ***
|Index equation for HOSPITAL
Constant| ***
AGE| ***
FEMALE| ***
HHNINC|
HHKIDS|
|Disturbance correlation
RHO(1,2)| ***

63. Marginal Effects What are the marginal effects Possible margins?
Effect of what on what?
Two equation model, what is the conditional mean?
Possible margins?
Derivatives of joint probability = Φ2(β1’xi1, β2’xi2,ρ)
Partials of E[yij|xij] =Φ(βj’xij) (Univariate probability)
Partials of E[yi1|xi1,xi2,yi2=1] = P(yi1,yi2=1)/Prob[yi2=1]
Note marginal effects involve both sets of regressors. If there are common variables, there are two effects in the derivative that are added.
(See Appendix for formulations.)

64. Marginal Effects: Decomposition

65. Direct Effects Derivatives of E[y1|x1,x2,y2=1] wrt x1
| Partial derivatives of E[y1|y2=1] with |
| respect to the vector of characteristics. |
| They are computed at the means of the Xs. |
| Effect shown is total of 4 parts above. |
| Estimate of E[y1|y2=1] = |
| Observations used for means are All Obs. |
| These are the direct marginal effects. |
|Variable | Coefficient | Standard Error |b/St.Er.|P[|Z|>z] | Mean of X|
AGE
FEMALE
EDUC
MARRIED
WORKING
HHNINC (Fixed Parameter)
HHKIDS (Fixed Parameter)

66. Indirect Effects Derivatives of E[y1|x1,x2,y2=1] wrt x2
| Partial derivatives of E[y1|y2=1] with |
| respect to the vector of characteristics. |
| They are computed at the means of the Xs. |
| Effect shown is total of 4 parts above. |
| Estimate of E[y1|y2=1] = |
| Observations used for means are All Obs. |
| These are the indirect marginal effects. |
|Variable | Coefficient | Standard Error |b/St.Er.|P[|Z|>z] | Mean of X|
AGE D
FEMALE
EDUC (Fixed Parameter)
MARRIED (Fixed Parameter)
WORKING (Fixed Parameter)
HHNINC
HHKIDS

67. Partial Effects: Total Effects Sum of Two Derivative Vectors
| Partial derivatives of E[y1|y2=1] with |
| respect to the vector of characteristics. |
| They are computed at the means of the Xs. |
| Effect shown is total of 4 parts above. |
| Estimate of E[y1|y2=1] = |
| Observations used for means are All Obs. |
| Total effects reported = direct+indirect. |
|Variable | Coefficient | Standard Error |b/St.Er.|P[|Z|>z] | Mean of X|
AGE
FEMALE
EDUC
MARRIED
WORKING
HHNINC
HHKIDS

68. Partial Effects: Dummy Variables Using Differences of Probabilities
| Analysis of dummy variables in the model. The effects are |
| computed using E[y1|y2=1,d=1] - E[y1|y2=1,d=0] where d is |
| the variable. Variances use the delta method. The effect |
| accounts for all appearances of the variable in the model.|
|Variable Effect Standard error t ratio
FEMALE
MARRIED
WORKING
HHKIDS

69. Average Partial Effects

70. Model Simulation

71. Model Simulation

72. A Simultaneous Equations Model

73. Fully Simultaneous “Model”
FIML Estimates of Bivariate Probit Model
Dependent variable DOCHOS
Log likelihood function
Variable| Coefficient Standard Error b/St.Er. P[|Z|>z] Mean of X
|Index equation for DOCTOR
Constant| ***
AGE| ***
FEMALE| ***
EDUC|
MARRIED|
WORKING|
HOSPITAL| ***
|Index equation for HOSPITAL
Constant| ***
AGE| ***
FEMALE| ***
HHNINC|
HHKIDS|
DOCTOR| ***
|Disturbance correlation
RHO(1,2)| *** ********

74. A Recursive Simultaneous Equations Model
Bivariate ; Lhs = y1,y2 ; Rh1=…,y2 ; Rh2 = … $

77. Causal Inference?

78. ATE and ATET for the RBP model

84. A Sample Selection Model

85. Sample Selection Model: Estimation

86. Application: Credit Scoring
American Express: 1992
N = 13,444 Applications
Observed application data
Observed acceptance/rejection of application
N1 = 10,499 Cardholders
Observed demographics and economic data
Observed default or not in first 12 months
Full Sample is in AmEx.lpj; description shows when imported.