# Simulation and Bayes Rule

## Bayesian Statistics Using Sampling Methods¶

This workbook adds more detail on the theoretical underpinnings of Metropolis Hastings MCMC and slightly tweaks and expounds on some examples from Thomas Wiecki's excellent blogpost on this topic. Of course, all errors are mine.

In order to understand our parameters $\theta$, MH MCMC trys to sample from our parameters' distribution, which is unknown. This distribution is the posterior, or $Pr(\theta|y)$. If we can somehow construct this distribution we can extract information about $\theta$.

To see this, consider the following trivial example. Suppose we have one parameter and are somehow able to construct the posterior distribution. For simplicity, suppose the posterior is distributed $N(0,1)$ (the standard normal distribution). If we take draws from this distribution, we can learn alot about its shape:

In [1]:
%matplotlib inline
import numpy as np
from scipy.stats import norm,uniform,lognorm
import matplotlib.pyplot as plt
import seaborn as sbn
from __future__ import division

sbn.set_style('white')
sbn.set_context('talk')

import warnings
warnings.filterwarnings('ignore')

/home/robhicks/anaconda/envs/econ400/lib/python2.7/site-packages/matplotlib/__init__.py:872: UserWarning: axes.color_cycle is deprecated and replaced with axes.prop_cycle; please use the latter.
warnings.warn(self.msg_depr % (key, alt_key))

In [2]:
# define the posterior distribution
def posterior(mean,std,N):
return norm.rvs(mean,std,N)

In [3]:
# take a random sample from the posterior:
sample = posterior(0,1,100000)

# calculate mean:
print "The mean of theta is ",np.mean(sample)
print "The standard deviation of theta is ", np.std(sample)
print "95% CI of theta is ", np.percentile(sample,[2.5,97.5])

# plot histogram
plt.title('Posterior Distribution of our Mean')
sbn.distplot(sample,bins=100)
plt.show()

The mean of theta is  0.0011068627216
The standard deviation of theta is  0.99509692537
95% CI of theta is  [-1.95834512  1.95521043]


Note, our sample values from the standard normal are exactly as one would expect and we hardly needed to sample from this posterior to know about $\theta$. For more interesting problems, we won't be able to construct (or know the properties of) the posterior as easily, so we need to devise a way to sample from the posterior to uncover information about $\theta$. This is exactly what the Metropolis Hastings Algorithm does.

### The Metropolis-Hastings Random Walk Algorithm¶

The Metropolis-Hastings algorithm is simply a random number generator from our posterior distribution.

The Metroplis-Hastings algorithm has been named one of the top 10 algorithms of the 20th century and appears on both the Math and Computational Sciences lists.

Consider constructing a set or series of potential beta vectors that explore the posterior distribution described above. When selecting beta vectors we would like to explore parameters having higher probability for the posterior ($Prob(\theta|\mathbf{X},\mathbf{Y},\alpha)$) . Essentially we want to construct a series of random draws from the posterior pdf of our parameters (called a chain) that is in regions of the parameter space that are high probability for the posterior. Once the chain is constructed, it reflects the underlying distribution of our parameter estimates.

#### Step 1. Generate proposal¶

We need to add another pdf to the mix: the proposal distribution. The proposal distribution is an exogenously assigned distribution from which we draw proposed values of our parameter vector $\theta$. Suppose we choose a symmetric proposal distribution such that $Prob(\theta_{t+1}|\theta_t) = Prob(\theta_t|\theta_{t+1})$. For example, we might specify that $\theta_{t+1} = \theta_t + N(0,\omega)$, where $N(0,\omega)$ is a random normal variate mean 0 standard deviation $\omega$. Both of the differences $(\theta_{t+1} - \theta_t)$ and $(\theta_{t+1} - \theta_t)$ are distributed normal with mean 0 and standard deviation $\omega$, so we say the proposal distribution is symmetric. Given our symmetric proposal distribution, denote the probability of $\theta^P_{t+1}$ conditional on $\theta_{t}$ as $Prob(\theta^P_{t+1}|\theta_t) = q(\theta^P_{t+1}|\theta_{t})$ Using a proposal distribution of this form leads to the random walk Metropolis-Hastings algorithm, the most common variety of MH in use.

Denote, $\theta_t$ as the current value of our Markov Chain and $\theta^P_{t+1}$ as the proposed value. Metropolis-Hastings lets us define an accept/reject criteria, where accept means that the candidate draw ($\theta_{t+1}$) provides information about $\theta$ in high probability regions, and is therefore a suitable draw from the posterior distribution of $\theta$.

The parameter $\omega$, the standard deviation of the random walk error term is often referred to as the sample width or proposal width parameter and we will turn back to this later in this notebook.

#### Step 2. Incorporate Bayes Rule¶

Recall that we can write Bayes Rule for our model as $$Prob(\theta |\mathbf{y}) = \frac{Prob(\mathbf{y}|\theta,\mathbf{x})Prob(\theta|\mathbf{x}))}{Prob(\mathbf{y}|\mathbf{x})}$$ Using Bayes Rule, we can evaluate how likely our proposed value ($\theta^P_{t+1}$) is relative to the current value in our chain ($\theta_{t+1}$):

$$\frac{Prob(\theta^P_{t+1}|\mathbf{y}) / q(\theta^P_{t+1}|\theta_t)}{Prob(\theta_{t}|\mathbf{y}) / q(\theta_t|\theta^P_{t+1})} = \frac{ \left ( \frac{Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x})Prob(\theta^P_{t+1}|\mathbf{x}))}{Prob(\mathbf{y}|\mathbf{x})} \right ) / q(\theta^P_{t+1}| \theta_t)} { \left ( \frac{Prob(\mathbf{y}|\theta_{t},\mathbf{x})Prob(\theta_{t}|\mathbf{x}))}{Prob(\mathbf{y}|\mathbf{x})} \right ) / q( \theta_t | \theta^P_{t+1})}$$

Noting that $q( \theta_t | \theta^P_{t+1}) = q(\theta^P_{t+1}| \theta_t)$ and $Prob(\mathbf{y}|\mathbf{x})$ is a normalizing constant irrespective of the value of $\theta$ being considered:

This complicated expression can be simplified to $$\frac{ Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x})Prob(\theta^P_{t+1}|\mathbf{x}))} { Prob(\mathbf{y}|\theta_{t},\mathbf{x})Prob(\theta_{t}|\mathbf{x}))}$$ this is a nice result, because we don't need to worry about calculating the probability of the evidence for the MH sampling method outlined below.

#### Step 3: Develop accept/reject criteria for $\theta^{P}_{t+1}$¶

Recall, that we want to sample from the posterior, so it should be the case that:

1. We won't likely include low probability proposals: $\theta^P_{t+1}$
2. We won't get stuck at very likely points like MAP

Now that we know the relative likelihood of the proposal value compared to the current value of $\theta$, Metropolis-Hastings showed that an acceptance criteria (sometimes called an accept/reject criteria) should explore the posterior distribution of $\theta$ by sampling points in proportion to how likely they are given the data. MH does this using the following rules:

1. If $\frac{ Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x})Prob(\theta^P_{t+1}|\mathbf{x}))} { Prob(\mathbf{y}|\theta_{t},\mathbf{x})Prob(\theta_{t}|\mathbf{x}))}$ > 1, then $\theta_{t+1} = \theta^P_{t+1}$
2. Else, $\theta_{t+1} = \theta^P_{t+1}$ if $u \le \frac{ Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x})Prob(\theta^P_{t+1}|\mathbf{x}))} { Prob(\mathbf{y}|\theta_{t},\mathbf{x})Prob(\theta_{t}|\mathbf{x}))}$
1. Else $\theta_{t+1} = \theta_t$

Notice condition (2) ensures that even inferior values of $\theta$ relative to the current value in the chain will sometimes be accepted in the chain. The acceptance rate for these values will be in proportion to their relative likelihood. The chain never converges, it dances around regions of the posterior where more likely values of $\theta$ reside. There is always a non-zero probability that the chain will deviate (and store values of the estimate of $\theta$ that actually cause the posterior to decline relative to the its value previously in the chain)!

Usually, the conditions outlined above are written in terms of the accept/reject criteria, $c$.
$$c(\theta^P_{t+1},\theta_t) = min \left( 1,\frac{ Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x})Prob(\theta^P_{t+1}|\mathbf{x}))} { Prob(\mathbf{y}|\theta_{t},\mathbf{x})Prob(\theta_{t}|\mathbf{x}))}\right )$$

Having calculated the accept/reject criteria value $c$, we need to decide whether to include $\theta^P_{t+1}$ in our chain. To do that, draw the random uniform value $u\in[0,1]$. If $c(\theta_{t+1},\theta_t)\ge u$ then add $\theta^P_{t+1}$ to our chain as $\theta_{t+1}$ that will help us characterize the distribution of parameters that most satisfy the posterior probability.

The Condition in Logs

For computational reasons, we usually work in logs, making these conditions: $$c(\theta^P_{t+1},\theta_t) = min \left( 0, log(Prob(\mathbf{y}|\theta^P_{t+1},\mathbf{x}))+log(Prob(\theta^P_{t+1}|\mathbf{x})))) - log( Prob(\mathbf{y}|\theta_{t},\mathbf{x})) - log(Prob(\theta_{t}|\mathbf{x}))) \right )$$ we accept $\theta^P_{t+1}$ when $c(\theta_{t+1},\theta_t)\ge log(u)$

##### Intuition of the MH Algorithm¶

Suppose a drunk lives up the hill from the bar. The drunk is unable to read roadsigns and only knows the house is at the very top of the hill. Like any drunk, this person stumbles and the intended direction (uphill) is usually (but not always) followed. By the time the drunk sobers up, the neighborhood uphill around his house has been thouroughly explored and in fact, the drunk has probably spent alot of time stumbling around in his own yard and even bumped into his front door many times without realizing he was home. The posterior is like the neighborhood uphill from the bar, the location of the house is the parameter representing the maximum posterior value, and the MH algorithm explores the posterior proportionally to the posterior probability values. So more time is spent in high likelihood areas.

#### A simple example¶

Suppose that your data is $y=[5,6]$ and you know that $\sigma_y = 1$. Using MH, let's construct a MCMC of the mean of $y$ of length 1 (not counting our starting value, which we will be arbitrarily set at $\mu_1 = 4$. Our priors are $\mu_0=6$ with standard deviation $\sigma_{\mu} = 1$.

In [4]:
# A simple example of the MH algorithm
y = np.array([5,6])      # data
std_y = 1  # known std dev of y

# priors
mu_0 = 6
std_mu_0 = 1

# proposal_width
omega=1.5

# arbitrary starting value of chain:
previous_val = 4

# notice the proposal distribution is normal, so
# we have the mh random walk as outlined above
mu_proposal = previous_val + omega*norm(0,1).rvs()

print "Previous Value=%2.3f and Proposal=%2.3f"%(previous_val,mu_proposal)

like_proposal = norm(mu_proposal,std_y).pdf(y).prod()
prior_proposal = norm(mu_0,std_mu_0).pdf(mu_proposal)

print "\nLikelihood at proposal: ", like_proposal
print "Prior at proposal:", prior_proposal

like_previous = norm(previous_val,std_y).pdf(y).prod()
prior_previous = norm(mu_0,std_mu_0).pdf(previous_val)

print "\nLikelihood at previous value: ", like_previous
print "Prior at previous value: ", prior_previous

c = np.min((1,like_proposal*prior_proposal/(like_previous*prior_previous)))

print "\nAccept Reject Criteria is: ", c
u = uniform.rvs()
print "Uniform Random Draw: ", u

current_value = (c>=u)*mu_proposal + (1-(c>=u))*previous_val
print "\nNext value in the chain is: %2.3f"% current_value

Previous Value=4.000 and Proposal=4.831

Likelihood at proposal:  0.0792703654705
Prior at proposal: 0.201545335128

Likelihood at previous value:  0.0130642332847
Prior at previous value:  0.0539909665132

Accept Reject Criteria is:  1.0
Uniform Random Draw:  0.825348358931

Next value in the chain is: 4.831

##### A Metropolis-Hastings Sampler¶

Rather than manually storing values, the following implements a loop that calculates a chain of samples from the posterior of desired length. Let's revisit our example from Analytical Bayes:

In [5]:
sigma = 3  # Note this is the std of data assumed known
data = norm(10,sigma).rvs(10)
mu_prior = 8
sigma_prior = 1  # Note this is our prior on the std of mu

In [6]:
ax = plt.subplot()
sbn.distplot(data, kde=False, ax=ax,bins=5)
_ = ax.set(title='Histogram of observed data', xlabel='x', ylabel='# observations');


Here are the likelihood and posterior analytical pdf's.

In [7]:
def calc_posterior_analytical(data, x, sigma, mu_0, sigma_0):
n = len(data)
# posterior parameter
mu_post = ( mu_0/(sigma_0**2) + data.sum()/(sigma**2) )*(1. / sigma_0**2 + n / sigma**2)**(-1)
sigma_post = np.sqrt((1. / sigma_0**2 + n / sigma**2)**(-1))
# probabilities
posterior = norm(mu_post,sigma_post).pdf(x)
prior = norm(mu_0,sigma_0).pdf(x)
return posterior,prior,mu_post,sigma_post


The function plot_proposal is rather long and cumbersome, but it is nice for generating plots that show how MH works iteration by iteration.

In [8]:
# allows us to plot what is happening inside the sampler iteration by iteration
def plot_proposal(mu_current, mu_proposal, sigma, mu_prior_mu, mu_prior_sd, data, accepted, trace, i):
from copy import copy
trace = copy(trace)
fig, (ax1, ax2, ax3, ax4) = plt.subplots(ncols=4, figsize=(16, 4))
fig.suptitle('Iteration %i' % (i + 1), fontsize=20,y=0.1)
x = np.linspace(3, 15, 1000)
color = 'g' if accepted else 'r'

# Plot prior
prior_current = norm(mu_prior_mu, mu_prior_sd).pdf(mu_current)
prior_proposal = norm(mu_prior_mu, mu_prior_sd).pdf(mu_proposal)
prior = norm(mu_prior_mu, mu_prior_sd).pdf(x)
ax1.plot(x, prior, color='k')
ax1.plot([mu_current] * 2, [0, prior_current], marker='o', color='b')
ax1.plot([mu_proposal] * 2, [0, prior_proposal], marker='o', color=color)
ax1.annotate("", xy=(mu_proposal, 0.2), xytext=(mu_current, 0.2),
arrowprops=dict(arrowstyle="->", lw=2.))
ax1.set(ylabel='Probability Density', title='current: prior(mu=%.2f) = %.2f\nproposal: prior(mu=%.2f) = %.2f' % (mu_current, prior_current, mu_proposal, prior_proposal))

# Likelihood
likelihood_current = norm(mu_current, sigma).pdf(data).prod()
likelihood_proposal = norm(mu_proposal, sigma).pdf(data).prod()
y = np.array([norm(loc=i,scale=sigma).pdf(data).prod() for i in x])
ax2.plot(x, y, color='k')
ax2.plot([mu_current,mu_current],[0,likelihood_current], color='b', marker='o', label='mu_current')
ax2.plot([mu_proposal,mu_proposal],[0,likelihood_proposal], color=color, marker='o', label='mu_proposal')
ax2.annotate("", xy=(mu_proposal, 0.8*1e-7), xytext=(mu_current, 0.8*1e-7),
arrowprops=dict(arrowstyle="->", lw=2.))
ax2.set(title='likelihood(mu=%.2f) = %.2e\nlikelihood(mu=%.2f) = %.2e' % (mu_current, likelihood_current, mu_proposal, likelihood_proposal))

# Posterior
posterior_analytical,prior,mu_post,sigma_post  = calc_posterior_analytical(data, x, sigma, mu_prior_mu, mu_prior_sd)
ax3.plot(x, posterior_analytical,color='k')
posterior_current, prior,mu_post,sigma_post  = calc_posterior_analytical(data, mu_current, sigma, mu_prior_mu, mu_prior_sd)
posterior_proposal,prior,mu_post,sigma_post  = calc_posterior_analytical(data, mu_proposal, sigma, mu_prior_mu, mu_prior_sd)
ax3.plot([mu_current] * 2, [0, posterior_current], marker='o', color='b')
ax3.plot([mu_proposal] * 2, [0, posterior_proposal], marker='o', color=color)
ax3.annotate("", xy=(mu_proposal, 0.2), xytext=(mu_current, 0.2),
arrowprops=dict(arrowstyle="->", lw=2.))
ax3.set(title='posterior(mu=%.2f) = %.5f\nposterior(mu=%.2f) = %.5f' % (mu_current, posterior_current, mu_proposal, posterior_proposal))

if accepted:
trace.append(mu_proposal)
else:
trace.append(mu_current)
ax4.plot(trace,color='k')
ax4.set(xlabel='iteration', ylabel='mu', title='trace')
plt.tight_layout()
plt.show()


The function sampler below is a simple Metropolis-Hastings sampler for 1 unknown parameter. It can spawn the plotter function to plot each iteration of the sampler, so beware if setting plot=True with large samples.

In [9]:
# a fairly basic mh mcmc random walk sampler:
def sampler(data, samples=4, mu_init=8, sigma= 1, proposal_width=3, plot=False,
mu_prior_mu=0, mu_prior_sd=1):
mu_current = mu_init
posterior = [mu_current]
for i in range(samples):
# suggest new position
mu_proposal = norm(mu_current, proposal_width).rvs()

# Compute likelihood by multiplying probabilities of each data point
likelihood_current = norm(mu_current, sigma).pdf(data).prod()
likelihood_proposal = norm(mu_proposal, sigma).pdf(data).prod()

# Compute prior probability of current and proposed mu
prior_current = norm(mu_prior_mu, mu_prior_sd).pdf(mu_current)
prior_proposal = norm(mu_prior_mu, mu_prior_sd).pdf(mu_proposal)

p_current = likelihood_current * prior_current
p_proposal = likelihood_proposal * prior_proposal

# Accept proposal?
c = np.min((1,p_proposal / p_current))

accept = np.random.rand() <= c

if plot:
plot_proposal(mu_current, mu_proposal, sigma, mu_prior_mu, mu_prior_sd, data, accept, posterior, i)

if accept:
# Update position
mu_current = mu_proposal

posterior.append(mu_current)

return np.array(posterior)

In [10]:
# the function needs to be called in order to produce something we can work with:
# here, we are going to only take 5 samples and plot what is happening.
# Note: green proposals are accepted, and red proposals are not accepted
chain_for_pic = sampler(data,samples=5,mu_init=8,sigma=sigma,proposal_width=1,plot=True,
mu_prior_mu=mu_prior,mu_prior_sd=sigma_prior)

In [11]:
# take lots of samples and turn plotting off.
chain = sampler(data,samples=5000,mu_init=0,sigma=3,proposal_width=3,plot=False,
mu_prior_mu=mu_prior,mu_prior_sd=sigma_prior)


The following plot, called a traceplot, is probably the first thing you should look at after the markov chain has been constructed:

In [12]:
plt.figure(figsize=(15,6))
plt.plot(chain[500:],lw=.8,alpha=.7)
plt.xlim(0,chain.shape[0]-500)
plt.xlabel('Sample')
plt.ylabel('$\\mu$')
plt.title('Traceplot for $\\mu$ (after 500 samples)')
plt.show()

In [13]:
plt.figure(figsize=(15,6))
plt.plot(chain[:500],lw=.9)
plt.xlim(0,500)
plt.xlabel('Sample')
plt.ylabel('$\\mu$')
plt.title('Traceplot for $\\mu$ (before 500 samples)')
plt.show()

In [14]:
plt.figure(figsize=(15,6))
plt.plot(chain[chain.shape[0]-500:-1],lw=.9)
plt.xlabel('Sample')
plt.ylabel('$\\mu$')
plt.title('Traceplot for $\\mu$ (last 500 samples)')
plt.show();

In [15]:
sbn.distplot(chain, kde=False, bins=50)
plt.xlabel('$\\mu$')
plt.show()

In [16]:
ax = plt.subplot()

sbn.distplot(chain[500:], ax=ax, label='simulated posterior')
x = np.linspace(5, 15, 500)
post,prior,mu_post,sigma_post = calc_posterior_analytical(data, x, sigma, mu_prior, sigma_prior)
ax.plot(x, post, 'g', label='analytic posterior')
_ = ax.set(xlabel='$\\mu$', ylabel='belief');
ax.legend();


### Practical issue #1: Proposal width¶

In all the analysis above, we assumed a proposal width of 3. What is the proposal width and why have we set it at 3? Recall that the proposal width $\omega$ is

$$\theta^P_{t+1} = \theta_t + N(0,\omega)$$

So larger values of the proposal width means that we are jumping further in the parameter space in our exploration of the posterior. Smaller values means we jump less.

Question: Will smaller values of the proposal width lead to lower or higher acceptance rates?

In [17]:
posterior_small = sampler(data,samples=5000,mu_init=6,sigma=3,proposal_width=.05,plot=False,
mu_prior_mu=mu_prior,mu_prior_sd=sigma_prior)
posterior_large = sampler(data,samples=5000,mu_init=6,sigma=3,proposal_width=9,plot=False,
mu_prior_mu=mu_prior,mu_prior_sd=sigma_prior)
posterior_medium = sampler(data,samples=5000,mu_init=6,sigma=3,proposal_width=3,plot=False,
mu_prior_mu=mu_prior,mu_prior_sd=sigma_prior)

In [18]:
plt.figure(figsize=(15,6))
plt.plot(posterior_small[0:],label='Proposal Width Small');
plt.plot(posterior_large[0:],label='Proposal Width Large',lw=1,alpha=.7)
plt.plot(posterior_medium[0:], label='Proposal Width Medium', lw=1, alpha=.7)
plt.xlabel('Sample')
plt.ylabel('$\\mu$')
plt.legend()
plt.show()

In [19]:
plt.figure(figsize=(10,18))
plt.subplot(3,1,1)
plt.title('Proposal Widths and the Analytical Posterior')
sbn.distplot(posterior_small[1000:], label='Proposal Width Small',color='g')
plt.plot(x, post, 'k', label='analytic posterior')
plt.legend()

plt.subplot(3,1,2)
sbn.distplot(posterior_large[1000:], label='Proposal Width Large',color='b')
plt.plot(x, post, 'k', label='analytic posterior')
plt.legend()

plt.subplot(3,1,3)
sbn.distplot(posterior_medium[1000:], label='Proposal Width Medium',color='r')
plt.plot(x, post, 'k', label='analytic posterior')
plt.legend()
plt.show()