# Conditional Maximum Likelihood for the Rasch Model

Tip

This example is also available as a Jupyter notebook: $rasch.ipynb$

The Rasch model is used in psychometrics as a model for assessment data such as student responses to a standardized test. Let $X_{pi}$ be the response accuracy of student $p$ to item $i$ where $X_{pi}=1$ if the item was answered correctly and $X_{pi}=0$ otherwise for $p=1,\ldots,n$ and $i=1,\ldots,m$. The model for this accuracy is

where $\xi_p > 0$ the latent ability of person $p$ and $\epsilon_i > 0$ is the difficulty of item $i$.

We simulate data from this model:

Random.seed!(123)
n = 1000
m = 5
theta = randn(n)
delta = randn(m)
r = zeros(n)
s = zeros(m)

for i in 1:n
p = exp.(theta[i] .- delta) ./ (1.0 .+ exp.(theta[i] .- delta))
for j in 1:m
if rand() < p[j] ##correct
r[i] += 1
s[j] += 1
end
end
end
f = [sum(r.==j) for j in 1:m];


Since the number of parameters increases with sample size standard maximum likelihood will not provide us consistent estimates. Instead we consider the conditional likelihood. It can be shown that the Rasch model is an exponential family model and that the sum score $r_p = \sum_{i} x_{pi}$ is the sufficient statistic for $\xi_p$. If we condition on the sum score we should be able to eliminate $\xi_p$. Indeed, with a bit of algebra we can show

where $\gamma_r(\mathbf\epsilon)$ is the elementary symmetric function of order $r$

where the sum is over all possible answer configurations that give a sum score of $r$. Algorithms to efficiently compute $\gamma$ and its derivatives are available in the literature (see eg Baker (1996) for a review and Biscarri (2018) for a more modern approach)

function esf_sum!(S::AbstractArray{T,1}, x::AbstractArray{T,1}) where T <: Real
n = length(x)
fill!(S,zero(T))
S = one(T)
@inbounds for col in 1:n
for r in 1:col
row = col - r + 1
S[row+1] = S[row+1] + x[col] * S[row]
end
end
end

function esf_ext!(S::AbstractArray{T,1}, H::AbstractArray{T,3}, x::AbstractArray{T,1}) where T <: Real
n = length(x)
esf_sum!(S, x)
H[:,:,1] .= zero(T)
H[:,:,2] .= one(T)

@inbounds for i in 3:n+1
for j in 1:n
H[j,j,i] = S[i-1] - x[j] * H[j,j,i-1]
for k in j+1:n
H[k,j,i] = S[i-1] - ((x[j]+x[k])*H[k,j,i-1] + x[j]*x[k]*H[k,j,i-2])
H[j,k,i] = H[k,j,i]
end
end
end
end

esf_ext! (generic function with 1 method)


The objective function we want to minimize is the negative log conditional likelihood

ϵ = ones(Float64, m)
β0 = zeros(Float64, m)
last_β = fill(NaN, m)
S = zeros(Float64, m+1)
H = zeros(Float64, m, m, m+1)

function calculate_common!(x, last_x)
if x != last_x
copyto!(last_x, x)
ϵ .= exp.(-x)
esf_ext!(S, H, ϵ)
end
end
function neglogLC(β)
calculate_common!(β, last_β)
return -s'log.(ϵ) + f'log.(S[2:end])
end

neglogLC (generic function with 1 method)


Parameter estimation is usually performed with respect to the unconstrained parameter $\beta_i = -\log{\epsilon_i}$. Taking the derivative with respect to $\beta_i$ (and applying the chain rule) one obtains

where $\gamma_{r-1}^{(i)} = \partial \gamma_{r}(\mathbf\epsilon)/\partial\epsilon_i$.

function g!(storage, β)
calculate_common!(β, last_β)
for j in 1:m
storage[j] = s[j]
for l in 1:m
storage[j] -= ϵ[j] * f[l] * (H[j,j,l+1] / S[l+1])
end
end
end

g! (generic function with 1 method)


Similarly the Hessian matrix can be computed

$$\dfrac{\partial^2 \log L_C(\mathbf\epsilon|\mathbf{r})}{\partial \beta_i\partial\beta_j} =$$

where $\gamma_{r-2}^{(i,j)} = \partial^2 \gamma_{r}(\mathbf\epsilon)/\partial\epsilon_i\partial\epsilon_j$.

function h!(storage, β)
calculate_common!(β, last_β)
for j in 1:m
for k in 1:m
storage[k,j] = 0.0
for l in 1:m
if j == k
storage[j,j] += f[l] * (ϵ[j]*H[j,j,l+1] / S[l+1]) *
(1 - ϵ[j]*H[j,j,l+1] / S[l+1])
elseif k > j
storage[k,j] += ϵ[j] * ϵ[k] * f[l] *
((H[k,j,l] / S[l+1]) - (H[j,j,l+1] * H[k,k,l+1]) / S[l+1] ^ 2)
else #k < j
storage[k,j] += ϵ[j] * ϵ[k] * f[l] *
((H[j,k,l] / S[l+1]) - (H[j,j,l+1] * H[k,k,l+1]) / S[l+1] ^ 2)
end
end
end
end
end

h! (generic function with 1 method)


The estimates of the item parameters are then obtained via standard optimization algorithms (either Newton-Raphson or L-BFGS). One last issue is that the model is not identifiable (multiplying the $\xi_p$ by a constant and dividing the $\epsilon_i$ by the same constant results in the same likelihood). Therefore some kind of constraint must be imposed when estimating the parameters. Typically either $\epsilon_1 = 0$ or $\prod_{i=1}^m \epsilon_i = 1$ (which is equivalent to $\sum_{i=1}^m \beta_i = 0$).

con_c!(c, x) = (c = sum(x); c)
function con_jacobian!(J, x)
J[1,:] .= ones(length(x))
end
function con_h!(h, x, λ)
for i in 1:size(h)
for j in 1:size(h)
h[i,j] += (i == j) ? λ : 0.0
end
end
end
lx = Float64[]; ux = Float64[]
lc = [0.0]; uc = [0.0]
df = TwiceDifferentiable(neglogLC, g!, h!, β0)
dfc = TwiceDifferentiableConstraints(con_c!, con_jacobian!, con_h!, lx, ux, lc, uc)
res = optimize(df, dfc, β0, IPNewton())

 * Status: success

* Candidate solution
Minimizer: [1.48e+00, 8.80e-01, -9.81e-01,  ...]
Minimum:   1.302751e+03

* Found with
Algorithm:     Interior Point Newton
Initial Point: [0.00e+00, 0.00e+00, 0.00e+00,  ...]

* Convergence measures
|x - x'|               = 2.67e-10 ≰ 0.0e+00
|x - x'|/|x'|          = 1.80e-10 ≰ 0.0e+00
|f(x) - f(x')|         = 0.00e+00 ≤ 0.0e+00
|f(x) - f(x')|/|f(x')| = 0.00e+00 ≤ 0.0e+00
|g(x)|                 = 4.67e-06 ≰ 1.0e-08

* Work counters
Seconds run:   0  (vs limit Inf)
Iterations:    26
f(x) calls:    61
∇f(x) calls:   61


Compare the estimate to the truth

delta_hat = res.minimizer
[delta delta_hat]

5×2 Array{Float64,2}:
1.14112    1.48015
0.597106   0.880231
-1.30405   -0.981096
-1.2566    -0.955468
-0.706518  -0.423819