© Copyright Quantopian Inc.
© Modifications Copyright QuantRocket LLC
Licensed under the Creative Commons Attribution 4.0.
© Copyright Quantopian Inc.
© Modifications Copyright QuantRocket LLC
Licensed under the Creative Commons Attribution 4.0.
By Delaney Granizo-Mackenzie and Andrei Kirilenko. Developed as part of the Masters of Finance curriculum at MIT Sloan.
AutoRegressive Conditionally Heteroskedastic (ARCH) occurs when the volatility of a time series is also autoregressive.
from functools import partial
import math
import numpy as np
import scipy
from scipy import stats
import statsmodels.api as sm
from statsmodels.stats.stattools import jarque_bera
import matplotlib.pyplot as plt
We'll start by using Monte Carlo sampling to simulate a GARCH(1, 1) process. Our dynamics will be
$$\sigma_1 = \sqrt{\frac{a_0}{1-a_1-b_1}} \\ \sigma_t^2 = a_0 + a_1 x_{t-1}^2+b_1 \sigma_{t-1}^2 \\ x_t = \sigma_t \epsilon_t \\ \epsilon \sim \mathcal{N}(0, 1)$$Our parameters will be $a_0 = 1$, $a_1=0.1$, and $b_1=0.8$. We will drop the first 10% (burn-in) of our simulated values.
# Define parameters
a0 = 1.0
a1 = 0.1
b1 = 0.8
sigma1 = math.sqrt(a0 / (1 - a1 - b1))
def simulate_GARCH(T, a0, a1, b1, sigma1):
# Initialize our values
X = np.ndarray(T)
sigma = np.ndarray(T)
sigma[0] = sigma1
for t in range(1, T):
# Draw the next x_t
X[t - 1] = sigma[t - 1] * np.random.normal(0, 1)
# Draw the next sigma_t
sigma[t] = math.sqrt(a0 + b1 * sigma[t - 1]**2 + a1 * X[t - 1]**2)
X[T - 1] = sigma[T - 1] * np.random.normal(0, 1)
return X, sigma
Now we'll compare the tails of the GARCH(1, 1) process with normally distributed values. We expect to see fatter tails, as the GARCH(1, 1) process will experience extreme values more often.
X, _ = simulate_GARCH(10000, a0, a1, b1, sigma1)
X = X[1000:] # Drop burn in
X = X / np.std(X) # Normalize X
def compare_tails_to_normal(X):
# Define matrix to store comparisons
A = np.zeros((2,4))
for k in range(4):
A[0, k] = len(X[X > (k + 1)]) / float(len(X)) # Estimate tails of X
A[1, k] = 1 - stats.norm.cdf(k + 1) # Compare to Gaussian distribution
return A
compare_tails_to_normal(X)
array([[1.58888889e-01, 2.48888889e-02, 2.22222222e-03, 0.00000000e+00],
[1.58655254e-01, 2.27501319e-02, 1.34989803e-03, 3.16712418e-05]])Sure enough, the tails of the GARCH(1, 1) process are fatter. We can also look at this graphically, although it's a little tricky to see.
plt.hist(X, bins=50)
plt.xlabel('sigma')
plt.ylabel('observations');
# Sample values from a normal distribution
X2 = np.random.normal(0, 1, 9000)
both = np.matrix([X, X2])
# Plot both the GARCH and normal values
plt.plot(both.T, alpha=.7);
plt.axhline(X2.std(), color='yellow', linestyle='--')
plt.axhline(-X2.std(), color='yellow', linestyle='--')
plt.axhline(3*X2.std(), color='red', linestyle='--')
plt.axhline(-3*X2.std(), color='red', linestyle='--')
plt.xlabel('time')
plt.ylabel('sigma');
What we're looking at here is the GARCH process in blue and the normal process in green. The 1 and 3 std bars are drawn on the plot. We can see that the blue GARCH process tends to cross the 3 std bar much more often than the green normal one.
The first step is to test for ARCH conditions. To do this we run a regression on $x_t$ fitting the following model.
$$x_t^2 = a_0 + a_1 x_{t-1}^2 + \dots + a_p x_{t-p}^2$$We use OLS to estimate $\hat\theta = (\hat a_0, \hat a_1, \dots, \hat a_p)$ and the covariance matrix $\hat\Omega$. We can then compute the test statistic
$$F = \hat\theta \hat\Omega^{-1} \hat\theta'$$We will reject if $F$ is greater than the 95% confidence bars in the $\mathcal(X)^2(p)$ distribution.
To test, we'll set $p=20$ and see what we get.
X, _ = simulate_GARCH(1100, a0, a1, b1, sigma1)
X = X[100:] # Drop burn in
p = 20
# Drop the first 20 so we have a lag of p's
Y2 = (X**2)[p:]
X2 = np.ndarray((980, p))
for i in range(p, 1000):
X2[i - p, :] = np.asarray((X**2)[i-p:i])[::-1]
model = sm.OLS(Y2, X2)
model = model.fit()
theta = np.matrix(model.params)
omega = np.matrix(model.cov_HC0)
F = (theta * np.linalg.inv(omega) * theta.T).item()
print(np.asarray(theta.T).shape)
plt.plot(range(20), np.asarray(theta.T))
plt.xlabel('Lag Amount')
plt.ylabel('Estimated Coefficient for Lagged Datapoint')
print('F = ' + str(F))
chi2dist = scipy.stats.chi2(p)
pvalue = 1-chi2dist.cdf(F)
print('p-value = ' + str(pvalue))
# Finally let's look at the significance of each a_p as measured by the standard deviations away from 0
print(theta/np.diag(omega))
(20, 1) F = 372.536098862269 p-value = 0.0 [[ 45.15349371 51.06809239 56.85108307 18.09338055 -14.37066069 14.9677237 18.60079392 73.92456045 53.07609156 21.58535032 -71.8879898 16.04688355 54.29480173 70.36315946 -51.61749873 38.60044005 -8.46430263 23.59115112 -12.62176926 27.84108897]]
Once we've decided that the data might have an underlying GARCH(1, 1) model, we would like to fit GARCH(1, 1) to the data by estimating parameters.
To do this we need the log-likelihood function
$$\mathcal{L}(\theta) = \sum_{t=1}^T - \ln \sqrt{2\pi} - \frac{x_t^2}{2\sigma_t^2} - \frac{1}{2}\ln(\sigma_t^2)$$To evaluate this function we need $x_t$ and $\sigma_t$ for $1 \leq t \leq T$. We have $x_t$, but we need to compute $\sigma_t$. To do this we need to make a guess for $\sigma_1$. Our guess will be $\sigma_1^2 = \hat E[x_t^2]$. Once we have our initial guess we compute the rest of the $\sigma$'s using the equation
$$\sigma_t^2 = a_0 + a_1 x_{t-1}^2 + b_1\sigma_{t-1}^2$$X, _ = simulate_GARCH(10000, a0, a1, b1, sigma1)
X = X[1000:] # Drop burn in
# Here's our function to compute the sigmas given the initial guess
def compute_squared_sigmas(X, initial_sigma, theta):
a0 = theta[0]
a1 = theta[1]
b1 = theta[2]
T = len(X)
sigma2 = np.ndarray(T)
sigma2[0] = initial_sigma ** 2
for t in range(1, T):
# Here's where we apply the equation
sigma2[t] = a0 + a1 * X[t-1]**2 + b1 * sigma2[t-1]
return sigma2
Let's look at the sigmas we just generated.
plt.plot(range(len(X)), compute_squared_sigmas(X, np.sqrt(np.mean(X**2)), (1, 0.5, 0.5)))
plt.xlabel('Time')
plt.ylabel('Sigma');
Now that we can compute the $\sigma_t$'s, we'll define the actual log likelihood function. This function will take as input our observations $x$ and $\theta$ and return $-\mathcal{L}(\theta)$. It is important to note that we return the negative log likelihood, as this way our numerical optimizer can minimize the function while maximizing the log likelihood.
Note that we are constantly re-computing the $\sigma_t$'s in this function.
def negative_log_likelihood(X, theta):
T = len(X)
# Estimate initial sigma squared
initial_sigma = np.sqrt(np.mean(X ** 2))
# Generate the squared sigma values
sigma2 = compute_squared_sigmas(X, initial_sigma, theta)
# Now actually compute
return -sum(
[-np.log(np.sqrt(2.0 * np.pi)) -
(X[t] ** 2) / (2.0 * sigma2[t]) -
0.5 * np.log(sigma2[t]) for
t in range(T)]
)
Now we perform numerical optimization to find our estimate for $$\hat\theta = \arg \max_{(a_0, a_1, b_1)}\mathcal{L}(\theta) = \arg \min_{(a_0, a_1, b_1)}-\mathcal{L}(\theta)$$
We have some constraints on this
$$a_1 \geq 0, b_1 \geq 0, a_1+b_1 < 1$$# Make our objective function by plugging X into our log likelihood function
objective = partial(negative_log_likelihood, X)
# Define the constraints for our minimizer
def constraint1(theta):
return np.array([1 - (theta[1] + theta[2])])
def constraint2(theta):
return np.array([theta[1]])
def constraint3(theta):
return np.array([theta[2]])
cons = ({'type': 'ineq', 'fun': constraint1},
{'type': 'ineq', 'fun': constraint2},
{'type': 'ineq', 'fun': constraint3})
# Actually do the minimization
result = scipy.optimize.minimize(objective, (1, 0.5, 0.5),
method='SLSQP',
constraints = cons)
theta_mle = result.x
print('theta MLE: ' + str(theta_mle))
<ipython-input-12-aebc6fba023b>:15: RuntimeWarning: invalid value encountered in log 0.5 * np.log(sigma2[t]) for
theta MLE: [1.05725561 0.09342784 0.80145309]
Now we would like a way to check our estimate. We'll look at two things:
We'll do both in our check_theta_estimate function.
def check_theta_estimate(X, theta_estimate):
initial_sigma = np.sqrt(np.mean(X ** 2))
sigma = np.sqrt(compute_squared_sigmas(X, initial_sigma, theta_estimate))
epsilon = X / sigma
print('Tails table')
print(compare_tails_to_normal(epsilon / np.std(epsilon)))
_, pvalue, _, _ = jarque_bera(epsilon)
print('Jarque-Bera probability normal: ' + str(pvalue))
check_theta_estimate(X, theta_mle)
Tails table [[1.57222222e-01 2.34444444e-02 1.33333333e-03 0.00000000e+00] [1.58655254e-01 2.27501319e-02 1.34989803e-03 3.16712418e-05]] Jarque-Bera probability normal: 0.8633608698256775
We've just computed an estimate using MLE, but we can also use Generalized Method of Moments (GMM) to estimate the GARCH(1, 1) parameters.
To do this we need to define our moments. We'll use 4.
# The n-th standardized moment
# skewness is 3, kurtosis is 4
def standardized_moment(x, mu, sigma, n):
return ((x - mu) ** n) / (sigma ** n)
GMM now has three steps.
Start with $W$ as the identity matrix.
Initialize $W$ and $T$ and define the objective function we need to minimize.
def gmm_objective(X, W, theta):
# Compute the residuals for X and theta
initial_sigma = np.sqrt(np.mean(X ** 2))
sigma = np.sqrt(compute_squared_sigmas(X, initial_sigma, theta))
e = X / sigma
# Compute the mean moments
m1 = np.mean(e)
m2 = np.mean(e ** 2) - 1
m3 = np.mean(standardized_moment(e, np.mean(e), np.std(e), 3))
m4 = np.mean(standardized_moment(e, np.mean(e), np.std(e), 4) - 3)
G = np.matrix([m1, m2, m3, m4]).T
return (G.T * W * G).item()
def gmm_variance(X, theta):
# Compute the residuals for X and theta
initial_sigma = np.sqrt(np.mean(X ** 2))
sigma = np.sqrt(compute_squared_sigmas(X, initial_sigma, theta))
e = X / sigma
# Compute the squared moments
m1 = e ** 2
m2 = (e ** 2 - 1) ** 2
m3 = standardized_moment(e, np.mean(e), np.std(e), 3) ** 2
m4 = (standardized_moment(e, np.mean(e), np.std(e), 4) - 3) ** 2
# Compute the covariance matrix g * g'
T = len(X)
s = np.ndarray((4, 1))
for t in range(T):
G = np.matrix([m1[t], m2[t], m3[t], m4[t]]).T
s = s + G * G.T
return s / T
Now we're ready to the do the iterated minimization step.
# Initialize GMM parameters
W = np.identity(4)
gmm_iterations = 10
# First guess
theta_gmm_estimate = theta_mle
# Perform iterated GMM
for i in range(gmm_iterations):
# Estimate new theta
objective = partial(gmm_objective, X, W)
result = scipy.optimize.minimize(objective, theta_gmm_estimate, constraints=cons)
theta_gmm_estimate = result.x
print('Iteration ' + str(i) + ' theta: ' + str(theta_gmm_estimate))
# Recompute W
W = np.linalg.inv(gmm_variance(X, theta_gmm_estimate))
check_theta_estimate(X, theta_gmm_estimate)
Iteration 0 theta: [1.05725167 0.0933636 0.80139481] Iteration 1 theta: [1.05725719 0.09341073 0.80144771] Iteration 2 theta: [1.05725719 0.09341073 0.80144771] Iteration 3 theta: [1.05725719 0.09341073 0.80144771] Iteration 4 theta: [1.05725719 0.09341073 0.80144771] Iteration 5 theta: [1.05725719 0.09341073 0.80144771] Iteration 6 theta: [1.05725719 0.09341073 0.80144771] Iteration 7 theta: [1.05725719 0.09341073 0.80144771] Iteration 8 theta: [1.05725719 0.09341073 0.80144771] Iteration 9 theta: [1.05725719 0.09341073 0.80144771] Tails table [[1.57222222e-01 2.34444444e-02 1.33333333e-03 0.00000000e+00] [1.58655254e-01 2.27501319e-02 1.34989803e-03 3.16712418e-05]] Jarque-Bera probability normal: 0.8633658949282679
Now that we've fitted a model to our observations, we'd like to be able to predict what the future volatility will look like. To do this, we can just simulate more values using our original GARCH dynamics and the estimated parameters.
The first thing we'll do is compute an initial $\sigma_t$. We'll compute our squared sigmas and take the last one.
sigma_hats = np.sqrt(compute_squared_sigmas(X, np.sqrt(np.mean(X**2)), theta_mle))
initial_sigma = sigma_hats[-1]
initial_sigma
2.625374960309133
Now we'll just sample values walking forward.
a0_estimate = theta_gmm_estimate[0]
a1_estimate = theta_gmm_estimate[1]
b1_estimate = theta_gmm_estimate[2]
X_forecast, sigma_forecast = simulate_GARCH(100, a0_estimate, a1_estimate, b1_estimate, initial_sigma)
plt.plot(range(-100, 0), X[-100:], 'b-')
plt.plot(range(-100, 0), sigma_hats[-100:], 'r-')
plt.plot(range(0, 100), X_forecast, 'b--')
plt.plot(range(0, 100), sigma_forecast, 'r--')
plt.xlabel('Time')
plt.legend(['X', 'sigma']);
One should note that because we are moving foward using a random walk, this analysis is supposed to give us a sense of the magnitude of sigma and therefore the risk we could face. It is not supposed to accurately model future values of X. In practice you would probably want to use Monte Carlo sampling to generate thousands of future scenarios, and then look at the potential range of outputs. We'll try that now. Keep in mind that this is a fairly simplistic way of doing this analysis, and that better techniques, such as Bayesian cones, exist.
from IPython.display import clear_output
plt.plot(range(-100, 0), X[-100:], 'b-')
plt.plot(range(-100, 0), sigma_hats[-100:], 'r-')
plt.xlabel('Time')
plt.legend(['X', 'sigma'])
max_X = [-np.inf]
min_X = [np.inf]
for i in range(100):
print(f'iteration {i+1} of 100')
clear_output(wait=True)
X_forecast, sigma_forecast = simulate_GARCH(100, a0_estimate, a1_estimate, b1_estimate, initial_sigma)
if max(X_forecast) > max(max_X):
max_X = X_forecast
elif min(X_forecast) < min(max_X):
min_X = X_forecast
plt.plot(range(0, 100), X_forecast, 'b--', alpha=0.05)
plt.plot(range(0, 100), sigma_forecast, 'r--', alpha=0.05)
# Draw the most extreme X values specially
plt.plot(range(0, 100), max_X, 'g--', alpha=1.0)
plt.plot(range(0, 100), min_X, 'g--', alpha=1.0);
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