# Modeling the Time Evolution of the Annualized Rate of Public Mass Shootings with Gaussian Processes¶

Nathan Sanders, Victor Lei (Legendary Entertainment)

January, 2017

## Abstract¶

Much of the public policy debate over gun control and gun rights in the United States hinges on the alarming incidence of public mass shootings, here defined as attacks killing four or more victims. Several times in recent years, individual, highly salient public mass shooting incidents have galvanized public discussion of reform efforts. But deliberative legislative action proceeds over a much longer timescale that should be informed by knowledge of the long term evolution of these events. We have used Stan to develop a new model for the annualized rate of public mass shootings in the United States based on a Gaussian process with a time-varying mean function. This design yields a predictive model with the full non-parametric flexibility of a Gaussian process, while retaining the direct interpretability of a parametric model for long-term evolution of the mass shooting rate. We apply this model to the Mother Jones database of public mass shootings and explore the posterior consequences of different prior choices and of correlations between hyperparameters. We reach conclusions about the long term evolution of the rate of public mass shootings in the United States and short-term periods deviating from this trend.

## Background¶

Tragic, high profile public events over the past few years like the shootings at the Washington Navy Yard; the Emanuel AME Church in Charleston; San Bernadino, CA; and Orlando, FL have raised public awareness of the dangers posed by public mass shooting events and sociological interest in understanding the motivations and occurrence rates of such events. There is no commonly accepted definition of a public mass shooting, but such an event is generally understood to be the simultaneous homicide of multiple people perpetrated by an individual or coordinated group via firearm.

A particular question facing elevated public, political, and scholarly scrutiny is whether the rate of public mass shootings has increased significantly over recent years. Lott (2014) responded to a September, 2013 FBI report on public mass shootings by re-evaluating sources of bias, reviewing data consistency, and redefining the period under consideration to conclude that no statistically significant increase is identifiable. Lott's work has been the subject of persistent controversy (see e.g. Johnson et al. 2012). In contrast, Cohen et al. (2014) claim that the rate of public mass shootings tripled over the four year period 2011-2014 based on a Statistical Process Control (SPC) analysis of the duration between successive events.

In this study, we present a new statistical approach to evaluating the time evolution of the rate of public mass shootings. We do not present original data on occurrences in the United States, address the myriad considerations inherent in defining a "mass shooting" event, or seek to resolve the causal issues of why the growth rate may have changed over time. We do adopt a commonly cited public mass shooting dataset and definition from Mother Jones.

We develop a Gaussian process-based model for the time evolution of the occurrence rate of public mass shootings and demonstrate inference under this model by straightforward application of the probabilistic programming language Stan. We use this case to explore the intersection of parametric and non-parametric models. We seek to merge a parametric model, with straightforward interpretations of posterior marginalized parameter inferences, with a non-parametric model that captures and permits discovery of unspecified trends. Stan's flexible modeling language permits rapid model design and iteration, while the No-U-Turn sampler allows us to fully explore the model posterior and understand the dependence between the parametric and non-parametric components of our model and the implications of our prior assumptions.

In the following notebook, we describe the Mother Jones dataset on US public mass shootings and lay out our statistical model and inference scheme. We then discuss the results from this inference, how they depend on choices for the prior distribution, and explore correlations between hyperparameters. Finally, we discuss the conclusions that can be reached from inspection of the marginal posterior distributions.

In [1]:
## Notebook setup
%matplotlib inline

import pandas as pd
import numpy as np
import pickle, os, copy
import scipy
from matplotlib import pyplot as plt
from matplotlib import cm
from matplotlib.ticker import FixedLocator, MaxNLocator, AutoMinorLocator

## NOTE: We encounter an error with this model using PyStan 2.14,
## so for now we will wrap cmdstan using stanhelper instead.
#import pystan

## See https://github.com/akucukelbir/stanhelper
import stanhelper
import subprocess
cmdstan_path = os.path.expanduser('~/Stan/cmdstan_2.14.0/')

from scipy import stats as sstats


### Package versions¶

In [2]:
%load_ext watermark
%watermark -v -m -p pandas,numpy,scipy,matplotlib,pystan

CPython 2.7.6
IPython 5.1.0

pandas 0.18.1
numpy 1.11.3
scipy 0.18.1
matplotlib 1.4.3
pystan 2.14.0.0

compiler   : GCC 4.8.4
system     : Linux
release    : 3.16.0-38-generic
machine    : x86_64
processor  : x86_64
CPU cores  : 4
interpreter: 64bit

In [3]:
print subprocess.check_output(cmdstan_path+'bin/stanc --version', shell=1)

stanc version 2.14.0



## Data¶

For this study, we consider the database published by Mother Jones (retrieved for this study on October 16, 2016; as of January 14, 2017, Mother Jones had not added any further events to its database for 2016), compiling incidents of public mass shootings in the United States from 1982 through the end of 2016. The database includes rich (quantitative and qualitative) metadata on the effects of the incidents, the mental health condition of the perpetrators, weapon type, how the perpetrators obtained their weapons, and more; however, we focus primarily on the dates of incident occurrence.

The definition of a public mass shooting is not universally agreed upon, and even when a firm definition is adopted there can be ambiguity in how to apply it to the complex and uncertain circumstances of these chaotic events. See Fox & Levin (2015) for a recent discussion. The criteria for inclusion in the Mother Jones database were described in a 2014 article by Mark Follman:

[The database] includes attacks in public places with four or more victims killed, a baseline established by the FBI a decade ago. We excluded mass murders in private homes related to domestic violence, as well as shootings tied to gang or other criminal activity.''

Follman discusses their motivations for these criteria and provide some examples of prominent incidents excluded by the criteria, such as the shooting at Ft. Hood in April, 2014. Note that the federal threshold for investigation of public mass shootings was lowered to three victim fatalities in January of 2013, and the Mother Jones database includes shootings under this more expansive definition starting from that date. To maintain a consistent definition for public mass shootings throughout the studied time period, we only consider shootings with four or more victim fatalities.

Our primary dataset is the count of incidents reported in this database per calendar year. We include incidents labeled as both "Mass" or "Spree" by Mother Jones.

In [4]:
## Load data

## Stadardize on definition of fatalities at 4.  Mother Jones changed it to 3 in 2013.
data = data[data.Fatalities > 3]

## Prepare data
# Aggregate data anually
data_annual = data.groupby('Year')
# Count cases by year and fill in empty years
cases_resamp = data_annual.count().Case.ix[np.arange(1982,2017)].fillna(0)
# Enumerate years in range
data_years = cases_resamp.index.values
# Enumerate quarters across daterange for later plotting
data_years_samp = np.arange(min(data_years), max(data_years)+10, .25)
# Format for Stan
stan_data = {
'N1': len(cases_resamp),
'x1': data_years - min(data_years),
'z1': cases_resamp.values.astype(int),
'N2': len(data_years_samp),
'x2': data_years_samp - min(data_years),
}

In [5]:
## Print the stan model inputs
for key in stan_data:
print key
print stan_data[key]
print '\n'

x2
[  0.     0.25   0.5    0.75   1.     1.25   1.5    1.75   2.     2.25
2.5    2.75   3.     3.25   3.5    3.75   4.     4.25   4.5    4.75   5.
5.25   5.5    5.75   6.     6.25   6.5    6.75   7.     7.25   7.5
7.75   8.     8.25   8.5    8.75   9.     9.25   9.5    9.75  10.    10.25
10.5   10.75  11.    11.25  11.5   11.75  12.    12.25  12.5   12.75  13.
13.25  13.5   13.75  14.    14.25  14.5   14.75  15.    15.25  15.5
15.75  16.    16.25  16.5   16.75  17.    17.25  17.5   17.75  18.    18.25
18.5   18.75  19.    19.25  19.5   19.75  20.    20.25  20.5   20.75  21.
21.25  21.5   21.75  22.    22.25  22.5   22.75  23.    23.25  23.5
23.75  24.    24.25  24.5   24.75  25.    25.25  25.5   25.75  26.    26.25
26.5   26.75  27.    27.25  27.5   27.75  28.    28.25  28.5   28.75  29.
29.25  29.5   29.75  30.    30.25  30.5   30.75  31.    31.25  31.5
31.75  32.    32.25  32.5   32.75  33.    33.25  33.5   33.75  34.    34.25
34.5   34.75  35.    35.25  35.5   35.75  36.    36.25  36.5   36.75  37.
37.25  37.5   37.75  38.    38.25  38.5   38.75  39.    39.25  39.5
39.75  40.    40.25  40.5   40.75  41.    41.25  41.5   41.75  42.    42.25
42.5   42.75  43.    43.25  43.5   43.75]

N1
35

N2
176

x1
[ 0  1  2  3  4  5  6  7  8  9 10 11 12 13 14 15 16 17 18 19 20 21 22 23 24
25 26 27 28 29 30 31 32 33 34]

z1
[1 0 2 0 1 1 1 2 1 3 2 4 1 1 1 2 3 5 1 1 0 1 1 2 3 4 3 4 1 3 7 5 3 4 4]


In [6]:
## Number of years with data
print len(stan_data['x1'])

35

In [7]:
## Number of interpolated points to do prediction for
print len(stan_data['x2'])

176


## Statistical Model¶

We adopt a univariate Gaussian process model (see e.g. Rasmussen & Williams 2006) as a non-parametric description of the time evolution of the annualized occurrence rate. The Gaussian process describes deviations from a mean function by a covariance matrix that controls the probability of the deviation as a function of the time differential between points. Robers et al. (2012) surveyed applications of Gaussian process models to timeseries data, and explored the implications of different choices for the mean and covariance functions.

We adopt the following system of units for the Gaussian Process model. The time vector $x$ is measured in years since 1982 and the outcome vector $z$ is defined as the number of occurrences per year.

Many applications of Gaussian processes adopt a constant, zero mean function. In that case, the relationship between the dependent variable(s) and the predictors is described entirely by the non-parametric family of functions generated from the Gaussian process covariance function.

We adopt a linear mean function and a squared-exponential covariance function. The mean function $\mu(x)$ is simply:

$$\mu(x) = \mu_0 + \mu_b~x$$

Note that we use a logarithmic parameterization of the likelihood for the occurence rate (see below), so the linear mean function corresponds to an exponential function for the evolution of the rate of shootings per year.

The familiar squared-exponential covariance function, which generates infinitely-differentiable functions from the Gaussian process, is:

$$k(x)_{i,j} = \eta^2~exp \big( -\rho^2 \sum_{d=1}^{D}(x_{i,d} - x_{j,d})^2 \big) + \delta_{i,j}~\sigma^2$$

where the hyperparameter $\eta$ controls the overall strength of covariance, $\rho$ controls the timescale over which functions drawn from the process vary, and $\sigma$ controls the baseline level of variance.

Our likelihood assumes that the occurrence rate is specified by exponentiated draws of the occurrence rate $y$ from the mean and covariance functions, and the observed outcome data is negative binomial-distributed according to the rate.

\begin{align} y(x) \sim \rm{N}(\mu(x), k(x)^2) \\ z(x) ~ \sim \rm{NB}(exp(y(x)), \phi) \end{align}

where $\rm{N}$ is the normal (parameterized by the standard deviation rather than the variance, per Stan standard syntax) and $\rm{NB}$ is the negative binomial distribution. We use the "alternative" parameterization of the negative binomial distribution described in the Stan manual, where the second parameter directly scales the overdispersion relative to a Poisson distribution. While we choose the negative binomial to permit overdispersion in the annualized mass shooting rate beyond counting noise, as we will see, the data provide strong evidence for small values of $\phi^{-1}$, consistent with Poisson noise.

The role of each component of the Gaussian process will depend largely on the timescale parameter $\rho$. When the timescale is short, the model effectively divides the response into a long-term (timescale of the range of the data; in this case, decades) parametric effect and a short-term (timescale of e.g. years) non-parametric effect. This approach gives us the full flexibility of the Gaussian process for predictive applications, while still allowing us to make interpretable, parametric inferences on the long-term evolution of the system.

We apply the following prior and hyperprior distributions to provide weak information about the scale of the relevant parameters in the adopted unit system:

\begin{align*} \rho^{-1} \sim \Gamma(\alpha_{\rho}, \beta_{\rho}) \\ \eta^2 \sim \rm{C}(2. 5) \\ \sigma^2 \sim \rm{C}(0, 2.5) \\ \mu_0 \sim \rm{N}(0, 2) \\ \mu_b \sim \rm{N}(0, 0.2) \\ \phi^{-1} \sim C(0, 5) \end{align*}

where $\Gamma$ is the gamma distribution; $\rm{C}$ is the half-Cauchy distribution; the parameters $\eta^2$, $\sigma^2$, and $\phi^{-1}$ are constrained to be positive; and we apply the constraint $\rho^{-1} > 1$ to enforce timescales $>1$ yr (the spacing of our data).

Below we explore different choices for the $\alpha$ and $\beta$ parameters of the gamma hyperprior on $\rho^{-1}$, labeled as $\alpha_{\rho}$ and $\beta_{\rho}$. In particular, we explore $(\alpha_{\rho},\beta_{\rho}) = (4,1)$ and $(1,1/100)$. These correspond to prior distributions with standard deviations of $2$ and $100$ years, respectively. On top of the linear trend in the mean function, the former represents a strong prior expectation that the annualized rate of public mass shootings evolves on a timescale of a few years, and the latter represents a nearly-flat expectation for variations on timescales from a few years to a few centuries.

We implement the Gaussian process model in Stan, adapting the logistic classification example in Section 14.5 of the Stan manual. Stan's NUTS sampler performs full joint Bayesian estimation of all parameters, including the mean function parameters $\mu_0$ and $\mu_b$ and the Gaussian Process hyperparmeters $\eta$, $\rho$, and $\sigma$ and the negative binomial over-dispersion $\phi^{-1}$. The $\alpha_{\rho}$ and $\beta_{\rho}$ hyperparameters of the $\rho$ hyperprior distribution are fixed. We use the Cholesky factor transformed implementation of the normal distribution to calculate the likelihood.

We expect these hyperparameters to be at least somewhat correlated and not well-identified, introducing significant curvature in the model posterior, indicating that Hamiltonian Monte Carlo (HMC) would be a particularly effective sampling strategy for this model (Betancourt & Girolami 2013). We fit the model to the 35 annual observations of the Mother Jones dataset and do model interpolation and prediction over a grid of 176 quarters from 1980 to 2024.

We typically fit 8 independent chains of length 2000 iterations (following an equal number of NUTS warmup samples) in parallel using Stan and observe a typical execution time of ~1 min. For the purposes of this notebook, we obtain a larger number of samples by fitting 20 chains of 4000 samples in order to improve the resolution of 2D posterior histograms.

In [8]:
with open('gp_model_final.stan', 'r') as f:
print stan_code

data {
int<lower=1> N1;
vector[N1] x1;
int z1[N1];
int<lower=1> N2;
vector[N2] x2;
real<lower=0> alpha_rho;
real<lower=0> beta_rho;
}
transformed data {
int<lower=1> N;
vector[N1+N2] x;
// cov_exp_quad wants real valued inputs
real rx[N1+N2];
real rx1[N1];
real rx2[N2];

N = N1 + N2;
x = append_row(x1, x2);

rx = to_array_1d(x);
rx1 = to_array_1d(x1);
rx2 = to_array_1d(x2);
}
parameters {
vector[N1] y_tilde1;
real<lower=0> eta_sq;
real<lower=1> inv_rho;
real<lower=0> sigma_sq;
real mu_0;
real mu_b;
real<lower=0> NB_phi_inv;
}
model {
vector[N1] mu1;
vector[N1] y1;
matrix[N1,N1] Sigma1;
matrix[N1,N1] L1;

// Calculate mean function
mu1 = mu_0 + mu_b * x1;

// GP hyperpriors
eta_sq ~ cauchy(0, 1);
sigma_sq ~ cauchy(0, 1);
inv_rho ~ gamma(alpha_rho, beta_rho); // Gamma prior with mean of 4 and std of 2

// Calculate covariance matrix using new optimized function
Sigma1 = cov_exp_quad(rx1, sqrt(eta_sq), sqrt(0.5) * inv_rho);
for (n in 1:N1) Sigma1[n,n] = Sigma1[n,n] + sigma_sq;

// Decompose
L1 = cholesky_decompose(Sigma1);
// We're using a the non-centered parameterization, so rescale y_tilde
y1 = mu1 + L1 * y_tilde1;

// Mean model priors
mu_0 ~ normal(0, 2);
mu_b ~ normal(0, 0.2);

// Negative-binomial prior
// For neg_binomial_2, phi^-1 controls the overdispersion.
// phi^-1 ~ 0 reduces to the poisson.  phi^-1 = 1 represents variance = mu+mu^2
NB_phi_inv ~ cauchy(0, 5);

// Generate non-centered parameterization
y_tilde1 ~ normal(0, 1);

// Likelihood
z1 ~ neg_binomial_2_log(y1, inv(NB_phi_inv));
}

generated quantities {
vector[N1] y1;
vector[N2] y2;
vector[N] y;
int z_rep[N];

{
// Don't save these parameters
matrix[N,N] Sigma;
matrix[N,N] L;
vector[N] y_tilde;

Sigma = cov_exp_quad(rx, sqrt(eta_sq), sqrt(0.5) * inv_rho);
for (n in 1:N) Sigma[n,n] = Sigma[n,n] + sigma_sq;

for (n in 1:N1) y_tilde[n] = y_tilde1[n];
for (n in (N1 + 1):N) y_tilde[n] = normal_rng(0,1);

// Decompose
L = cholesky_decompose(Sigma);
y = mu_0 + mu_b * x + L * y_tilde;

for (n in 1:N1) y1[n] = y[n];
for (n in 1:N2) y2[n] = y[N1+n];

for (n in 1:N) z_rep[n] = neg_binomial_2_log_rng(y[n], inv(NB_phi_inv));
}
}



Note that we use the newly introduced cov_exp_quad function to implement the squared exponential covariance function, and we rescale $\rho^{-1}$ by $2^{-1/2}$ to accomodate the difference between this implementation and our definition above. Moreover, we use a non-centered parameterization (see e.g. Papaspiliopoulos et al. 2003) for the Gaussian process, modeling the latent parameter $\tilde{y}$ as standard normal and then transforming to a sampled value for $y$ by rescaling by the covariance matrix.

## Model fitting¶

In [9]:
## Compile using pystan
#stan_model_compiled = pystan.StanModel(model_code=stan_code)

### Compile using cmdstan
### Script expects cmdstan installation at cmdstan_path
subprocess.call("mkdir "+cmdstan_path+"user-models", shell=1)
subprocess.call("cp gp_model_final.stan " + cmdstan_path+"user-models/", shell=1)
subprocess.call("make user-models/gp_model_final", cwd=cmdstan_path, shell=1)

Out[9]:
0

Below we explore the consequences of different choices for the prior distribution on $\rho^{-1}$. To facilitate that analysis, here we fit the model twice with two different hyperparameter specifications provided as data. We will visualize and discuss these hyperprior choices in the next section. When not explicitly making comparisons between the two models, we focus on the model with the stronger prior on $\rho^{-1}$.

In [10]:
## Sampling parameters
Nchains = 20
Niter = 8000
cdic = {'max_treedepth': 15, 'adapt_delta': 0.95}

In [11]:
## Sample with strong prior on rho
stan_data_rho_strong = copy.copy(stan_data)
stan_data_rho_strong['alpha_rho'] = 4
stan_data_rho_strong['beta_rho'] = 1

## Sample with pystan
#stan_model_samp_rho_strong = stan_model_compiled.sampling(
#    data = stan_data_rho_strong, iter=Niter,
#    chains=Nchains, control=cdic, seed=1002
#    )

## Sample with cmdstan
## Delete any old samples first
os.system('rm output_cmdstan_gp_rhostrong_samples*.csv')
stanhelper.stan_rdump(stan_data_rho_strong, 'input_data_rhostrong_final.R')
p = []
for i in range(Nchains):
cmd = """
{0}user-models/gp_model_final \
data file='input_data_rhostrong_final.R' \
sample num_warmup={2} num_samples={2} \
algorithm=hmc engine=nuts max_depth={3} \
random seed=1002 id={1} \
output file=output_cmdstan_gp_rhostrong_samples{1}.csv
p += [subprocess.Popen(cmd, shell=True)]

## Don't move on until sampling is complete.
for i in range(Nchains):
p[i].wait()

## Write out results if using pystan
#stan_model_ext_rho_strong = stan_model_samp_rho_strong.extract()
#with open('stan_model_ext_rho_strong.p','w') as f: pickle.dump(stan_model_ext_rho_strong,f)

In [12]:
## Sample with weak prior on rho
stan_data_rho_weak = copy.copy(stan_data)
stan_data_rho_weak['alpha_rho'] = 1
stan_data_rho_weak['beta_rho'] = 1/100.

## Sample with pystan
#stan_model_samp_rho_weak = stan_model_compiled.sampling(data = stan_data_rho_weak, iter=Niter, chains=Nchains, control=cdic)

## Sample with cmdstan
## Delete any old samples first
os.system('rm output_cmdstan_gp_rhoweak_samples*.csv')
stanhelper.stan_rdump(stan_data_rho_weak, 'input_data_rhoweak_final.R')
p = []
for i in range(Nchains):
cmd = """
{0}user-models/gp_model_final \
data file='input_data_rhoweak_final.R' \
sample num_warmup={2} num_samples={2} \
algorithm=hmc engine=nuts max_depth={3} \
random seed=1002 id={1} \
output file=output_cmdstan_gp_rhoweak_samples{1}.csv
p += [subprocess.Popen(cmd, shell=True)]

## Don't move on until sampling is complete.
for i in range(Nchains):
p[i].wait()

## Write out results if using pystan
#stan_model_ext_rho_weak = stan_model_samp_rho_weak.extract()
#with open('stan_model_ext_rho_weak.p','w') as f: pickle.dump(stan_model_ext_rho_weak,f)

In [13]:
def stan_read_csv_multi(path):
"""
from multiple chains.

Parameters:
* path: file path for cmdstan output files including wildcard (*)
"""
## Enumerate files
from glob import glob
files = glob(path)

result = {}
for file in files:

## Combine dictionaries
result_out = {}
keys = result[files[0]]
for key in keys:
result_out[key] = result[files[0]][key]
for f in files:
result_out[key] = np.append(result_out[key], result[f][key], axis=0)

## Remove extraneous dimension
for key in keys:
if result_out[key].shape[-1] == 1:
result_out[key] = np.squeeze(result_out[key], -1)

return result_out



The MCMC trace illustrates the high independence of samples achieved after the NUTS algorithm warm-up period, and the low variance in sampling distributions between chains.

In [14]:
## Traceplot
trace_pars = [('eta_sq','$\\eta^2$'),
('inv_rho','$\\rho^{-1}$'),
('sigma_sq','$\\sigma^2$'),
('mu_0','$\\mu_0$'),
('mu_b','$\\mu_b$'),
('NB_phi_inv','$\\rm{NB}_\\phi^{-1}$')]
fig,axs = plt.subplots(len(trace_pars),2, figsize=(8,8), sharex='all', sharey='row')
exts = [stan_model_ext_rho_strong, stan_model_ext_rho_weak]
exts_names = [r'Strong $\rho$ prior', r'Weak $\rho$ prior']
for j in range(2):
axs[0,j].set_title(exts_names[j])
for i,par in enumerate(trace_pars):
axs[i,j].plot(exts[j][par[0]], color='.5')
if j==0: axs[i,j].set_ylabel(par[1])
for k in range(1, Nchains+1):
axs[i,j].axvline(Niter/2 * k, c='r', zorder=-1)

axs[len(trace_pars) - 1,j].set_xticks(np.arange(0, (Niter/2)*Nchains+1, Niter*2))


We assess MCMC convergence quantitatively using the Gelman-Rubin convergence diagnostic, $\hat{R}$, a comparison of within- to between-chain variance. We find that $\hat{R} \ll 1.05$ for all parameters, indicating a negligable discrepancy in the sampling distributions between chains.

In [15]:
def read_stansummary(path, cmdstan_path=cmdstan_path):
"""
Wrapper for the cmdstan program stan_summary to calculate
sampling summary statistics across multiple MCMC chains.

Args:
path (str): Path, with a wildcard (*) for the id number
of each output chain

cmdstan_path (str): Path to the stan home directory

Returns:
out: A pandas dataframe with the summary statistics provided
by stan_summary.  Note that each element of array variables
are provided on separate lines
"""
from StringIO import StringIO
summary_string = subprocess.check_output(cmdstan_path + 'bin/stansummary --sig_figs=5 '+path, shell=1)
return out

## Use cmdstan's stansummary command to calculate rhat

In [16]:
## Get summary statistics using pystan
#model_summary = stan_model_samp_rho_strong.summary()
#Rhat_vec = model_summary['summary'][:,array(model_summary['summary_colnames'])=='Rhat']
#pars = model_summary['summary_rownames']

## Get summary statistics using cmdstan wrapper
model_summary = stan_model_sum_rho_strong
Rhat_vec = stan_model_sum_rho_strong['R_hat'].values
pars = stan_model_sum_rho_strong.index

## Replace y1, y2 with summaries
sel_pars = ['y1', 'y2', u'eta_sq', u'inv_rho', u'sigma_sq', u'mu_0', u'mu_b', 'NB_phi_inv']
Rhat_dic = {}
for spar in sel_pars:
if spar in ('y1','y2'):
sel = np.where([True if p.startswith(spar) else False for p in pars])
Rhat_dic[spar] = np.percentile(Rhat_vec[sel], [5,50,95])
else:
Rhat_dic[spar] = [Rhat_vec[[pars==spar]],]*3

plt.figure(figsize=(5,6))
plt.errorbar(np.array(Rhat_dic.values())[:,1], np.arange(len(sel_pars)), \
xerr= [np.array(Rhat_dic.values())[:,1] - np.array(Rhat_dic.values())[:,0],\
np.array(Rhat_dic.values())[:,2] - np.array(Rhat_dic.values())[:,1]],\
capsize=0, marker='o', color='k', lw=0)
plt.yticks(np.arange(len(sel_pars)), Rhat_dic.keys(), size=11)
plt.xlabel('$\hat{R}$')
plt.axvline(1.0, color='.5', ls='solid', zorder=-2)
plt.axvline(1.05, color='.5', ls='dashed', zorder=-2)
plt.ylim(-.5, len(sel_pars)-.5)
plt.xlim(0.99, 1.06)

Out[16]:
(0.99, 1.06)
/usr/local/lib/python2.7/dist-packages/matplotlib/collections.py:590: FutureWarning: elementwise comparison failed; returning scalar instead, but in the future will perform elementwise comparison
if self._edgecolors == str('face'):


## Posterior Simulations and Predictive Checks¶

To assess goodness of fit, we inspect simulated draws of the Gaussian process from the posterior and perform posterior predictive checks.

### Simulated draws¶

First we perform a posterior predictive check by visualizing the sampled values of $z$, which realizes both a draw from the latent Gaussian process for the public mass shootings rate and the overdispersed counting noise of the negative binomial distribution.

In [17]:
N_samp = Niter / 2
print len(stan_model_ext_rho_strong['z_rep'])
print Niter

fig, axs = plt.subplots(5,5, figsize=(7,7), sharex='all', sharey='all')
po = axs[0,0].plot(data_years, stan_data['z1'], 'o', c='k', mfc='k', label='Observations', zorder=2, lw=1, ms=4)
axs[0,0].legend(numpoints=1, prop={'size':6})
for i in range(1,25):
draw = np.random.randint(0, N_samp)
py = stan_model_ext_rho_strong['z_rep'][draw][:stan_data['N1']]
axs.flatten()[i].plot(data_years, py,  mfc='k', marker='o',
lw=.5, mec='none', ms=2, color='.5', label='GP realization')
axs[0,1].legend(numpoints=1, prop={'size':6})
axs[0,0].set_ylim(0,15)
axs[0,0].set_xticks([1980, 1990, 2000, 2010, 2020])
for ax in axs.flatten():
plt.setp(ax.get_xticklabels(), rotation='vertical', fontsize=9)
plt.setp(ax.get_yticklabels(), fontsize=9)

axs[2,0].set_ylabel('public mass shootings per year', size=9)

84000
8000

Out[17]:
<matplotlib.text.Text at 0x7f6557e22750>

Visual inspection suggests that the observations simulated under the model show similar variation over time as the actual observations (first panel). We note that some realizations have annual counts at the later end of the modeled time range that exceed the largest observed annual count (7 public mass shootings). Some exceedence is expected given the counting noise, but this posterior predictive check could guide revision of the prior on the over-dispersion parameter or the choice of the negative binomial likelihood.

Because the relative variance in the annualized counting statistics is high (i.e. public mass shootings are generally infrequent on an annual basis), it is also helpful to examine the model for the underlying shooting rate in detail. Next we plot the posterior distribution of the Gaussian process for the annualized mass shooting rate simulated across a grid of timepoints subsampled between years and extending beyond the current year (2016), effectively interpolating and extrapolating from the observations. The mean of the posterior predictive distribution of the Gaussian process is shown with the solid blue line, and the shaded region shows the 16 and 84th percentile intervals of the posterior (i.e. the "$1\sigma$ range").

In [18]:
def plot_GP(stan_model_ext):
y2_sum = np.percentile(np.exp(stan_model_ext['y2']), [16,50,84], axis=0)
plt.figure(figsize=(7,5))
pfb = plt.fill_between(data_years_samp, y2_sum[0], y2_sum[2], color='b', alpha=.5)
pfg = plt.plot(data_years_samp, y2_sum[1], c='b', lw=2, label='GP model', zorder=0)
po = plt.plot(data_years, stan_data['z1'], 'o', c='k', label='Observations', zorder=2)
plt.xlabel('Year')
plt.ylabel('Annual rate of public mass shootings')
plt.legend(prop={'size':10}, loc=2)
plt.ylim(0,15)
plt.gca().xaxis.set_minor_locator(FixedLocator(np.arange(min(data_years_samp), max(data_years_samp))))
plt.gca().set_xlim(min(data_years_samp) - 1, max(data_years_samp) + 1)
return pfb, pfg, po

pfb, pfg, po = plot_GP(stan_model_ext_rho_strong)


The Gaussian process captures an increase in the mass shooting rate over the decades and some fluctuations against that trend during certain periods, as we will explore in more detail below. The model does not show any visually apparent deviations from the evolution of the observational time series, although comparison to the data highlights several years with substantially outlying mass shooting totals (e.g. 1993 and 1999). The extrapolated period ($>2016$) suggests a range of possible future rates of growth from the 2016 level.

We add random draws from the mean function to visualize our inferences on the long-term time evolution of the mass shooting rate.

In [19]:
def plot_GP_mu_draws(stan_model_ext):
plot_GP(stan_model_ext)
N_samp = len(stan_model_ext['mu_0'])
px = np.linspace(min(data_years_samp), max(data_years_samp), 100)
pfms = []
for i in range(20):
draw = np.random.randint(0, N_samp)
py = np.exp(stan_model_ext['mu_0'][draw] + (px - min(data_years)) * stan_model_ext['mu_b'][draw])
pfms.append(plt.plot(px, py,  c='r',
zorder = 1, label = 'Mean function draws' if i==0 else None))
plt.legend(prop={'size':10}, loc=2)

plot_GP_mu_draws(stan_model_ext_rho_strong)


The comparison between draws of the mean functions (red) and the model posterior (blue) suggests that the mean function captures most of the modeled variation in the shooting rate over time.

We can understand the behavior of the Gaussian process covariance function by isolating it from the mean function. We do so by subtracting the linear component of the mean function from the simulated Gaussian process rates ($y_2$) and plotting against the observations.

In [20]:
y2_gp_rho_strong = np.percentile(np.exp(
stan_model_ext_rho_strong['y2'] -
np.dot(stan_model_ext_rho_strong['mu_b'][:,np.newaxis], (data_years_samp[np.newaxis,:] - min(data_years)))
), [16,25,50,75,84], axis=0)

fig, axs = plt.subplots(2, figsize=(7,7), sharex='all')
pfb = axs[1].fill_between(data_years_samp, y2_gp_rho_strong[1], y2_gp_rho_strong[3], color='b', alpha=.25)
pfb2 = axs[1].fill_between(data_years_samp, y2_gp_rho_strong[0], y2_gp_rho_strong[4], color='b', alpha=.25)
pfg = axs[1].plot(data_years_samp, y2_gp_rho_strong[2], c='b', lw=2, label='GP model (covariance only)', zorder=0)
po = axs[0].plot(data_years, stan_data['z1'], 'o', c='k', label='Observations', zorder=2)
axs[1].axhline(np.exp(stan_model_ext_rho_strong['mu_0'].mean()), color='orange', label='$\mu_0$')

axs[0].set_ylabel('Annual rate of \npublic mass shootings\n(observations)')
axs[1].legend(prop={'size':8}, loc=2, ncol=2)
axs[1].set_ylabel('Annual rate of \npublic mass shootings\n(model)')

axs[1].set_ylim(0, 2.2)
axs[1].xaxis.set_minor_locator(FixedLocator(np.arange(min(data_years_samp), max(data_years_samp))))
axs[1].set_xlim(min(data_years_samp) - 1, max(data_years_samp) + 1)

Out[20]:
(1981.0, 2026.75)

In this plot, the shaded regions show the interquartile and $[16-84]$th percentile ranges. The fact that the interquartile contours never cross the mean ($\mu_0$) indicates that there is never $>75\%$ probability that the annualized trend deviates from the linear mean function. However, there are times when the interquartile range approaches the mean.

Perhaps the most salient feature captured by the covariance function of the Gaussian process is a dip in the annualized rate of public mass shootings in the years from about 2000 to 2005. The model has no features that would seek to explain the causal origin of this dip, although many readers may be surprised by its juxtoposition with the Columbine High School massacre (1999), which is understood to have spawned dozens of "copycat" attacks over time (see e.g. Follman & Andrews 2015).

The largest positive deviation from the mean function occurs between about 1988 and 1993. During that time, the mean function itself is very small (see previous figure), so this does not reresent a large absolute deviation.

### Gaussian process with weak $\rho^{-1}$ prior¶

For comparison, we visualize the latent Gaussian process under a weak prior for $\rho^{-1}$.

In [21]:
plot_GP(stan_model_ext_rho_weak)

Out[21]:
(<matplotlib.collections.PolyCollection at 0x7f65619ea1d0>,
[<matplotlib.lines.Line2D at 0x7f6561d60750>],
[<matplotlib.lines.Line2D at 0x7f65614b6890>])

It's clear from this visualization that the Gaussian process does not capture significant short-timescale variations when the timescale prior is loosened. This model also generally expresses lower uncertainty in the annual public mass shootings rate. Consistent with the reliance on the parametric, linear mean function, the extrapolated predictions do not account for any substantial probability of decrease in the rate of public mass shootings after 2016.

We can see the dominance of the mean function over the covariance function directly by again visualizing the isolated Gaussian process covariance function, which shows virtually no deviation from the mean:

In [22]:
y2_gp_rho_weak = np.percentile(np.exp(
stan_model_ext_rho_weak['y2'] -
np.dot(stan_model_ext_rho_weak['mu_b'][:,np.newaxis], (data_years_samp[np.newaxis,:] - min(data_years)))
), [16,25,50,75,84], axis=0)

fig, axs = plt.subplots(1, figsize=(7,5), sharex='all')
pfb = axs.fill_between(data_years_samp, y2_gp_rho_weak[1], y2_gp_rho_weak[3], color='b', alpha=.25)
pfb2 = axs.fill_between(data_years_samp, y2_gp_rho_weak[0], y2_gp_rho_weak[4], color='b', alpha=.25)
pfg = axs.plot(data_years_samp, y2_gp_rho_weak[2], c='b', lw=2, label='GP model (covariance only)', zorder=0)
axs.axhline(np.exp(stan_model_ext_rho_weak['mu_0'].mean()), color='orange', label='$\mu_0$')

axs.legend(prop={'size':8}, loc=2, ncol=2)
axs.set_ylabel('Annual rate of \npublic mass shootings\n(model)')
axs.set_title(r'Weak $\rho$ prior')

axs.set_ylim(0, 2.2)
axs.xaxis.set_minor_locator(FixedLocator(np.arange(min(data_years_samp), max(data_years_samp))))
axs.set_xlim(min(data_years_samp) - 1, max(data_years_samp) + 1)

Out[22]:
(1981.0, 2026.75)

## Inspection of posterior correlations¶

Before we explore the marginalized posterior distributions of the parameters in our model, we take advantage of the fully Bayesian posterior samples generated by the NUTS simulations to understand the correlations between parameters in the posterior distribution.

First we note that the parameters of the linearized mean function are highly correlated:

In [23]:
plt.figure()
pa = plt.hist2d(stan_model_ext_rho_strong['mu_0'],
stan_model_ext_rho_strong['mu_b'],
bins=100, cmap=cm.Reds, cmin=4)
plt.xlabel(r'$\mu_0$ (log shootings)')
plt.ylabel(r'$\mu_b$ (log shootings per year)')
plt.axvline(0, color='k', ls='dashed')
plt.axhline(0, color='k', ls='dashed')
plt.axis([-1.5,1.5,-0.05,.1])
cb = plt.colorbar()
cb.set_label('Number of posterior samples')


If the mean rate of public mass shootings at the beginning of the time series ($\mu_0$) is inferred to be higher, then the increase in the mean function over time needed to explain the observations ($\mu_b$) would be lower. However, at all probable values of $\mu_0$, the distribution of $\mu_b$ is predominantly positive.

We can fit a simple linear model to understand more subtle correlations in the multivariate posterior distribution. Here we fit a model for $\rho^{-1}$ as a function of the other major parameters of the model. We standardize the predictors so that we can directly compare the coefficients on the linear model.

In [24]:
import statsmodels.api as sm

## Assemble data matrices
y = pd.Series(stan_model_ext_rho_strong['inv_rho']); y.name = 'inv_rho'
X = pd.DataFrame({
'eta':np.sqrt(stan_model_ext_rho_strong['eta_sq']),
'mu_0':stan_model_ext_rho_strong['mu_0'],
'mu_b':stan_model_ext_rho_strong['mu_b'],
'sigma':np.sqrt(stan_model_ext_rho_strong['sigma_sq']),
'NB_phi_inv':np.sqrt(stan_model_ext_rho_strong['NB_phi_inv']),
})
## Standardize
X = X - X.mean()
X = X / X.std()
y = (y - y.mean()) / y.std()
## Fit linear model using stats models
est = sm.OLS(y, X).fit()
## Print summary
print est.summary2()

                   Results: Ordinary least squares
====================================================================
Dependent Variable: inv_rho          AIC:                233702.4431
Date:               2017-01-14 23:07 BIC:                233758.4745
No. Observations:   84000            Log-Likelihood:     -1.1685e+05
Df Model:           5                F-statistic:        964.7
Df Residuals:       83994            Prob (F-statistic): 0.00
R-squared:          0.054            Scale:              0.94575
----------------------------------------------------------------------
Coef.    Std.Err.      t      P>|t|     [0.025   0.975]
----------------------------------------------------------------------
const         -0.0000     0.0034   -0.0000   1.0000   -0.0066   0.0066
NB_phi_inv     0.0170     0.0034    5.0545   0.0000    0.0104   0.0236
eta            0.2318     0.0034   68.7066   0.0000    0.2252   0.2384
mu_0           0.0585     0.0062    9.4253   0.0000    0.0463   0.0706
mu_b           0.0647     0.0062   10.4458   0.0000    0.0525   0.0768
sigma          0.0172     0.0034    5.0956   0.0000    0.0106   0.0238
--------------------------------------------------------------------
Omnibus:             11670.523      Durbin-Watson:         2.016
Prob(Omnibus):       0.000          Jarque-Bera (JB):      18349.721
Skew:                0.978          Prob(JB):              0.000
Kurtosis:            4.189          Condition No.:         3
====================================================================



We see that the most significant correlation is between $\rho^{-1}$ and $\eta$. When we visualize this correlation, we observe that the level of posterior curvature associated with these two variables is small, though significant.

In [25]:
plt.figure()
pa = plt.hist2d(np.sqrt(stan_model_ext_rho_strong['eta_sq']),
stan_model_ext_rho_strong['inv_rho'],
bins=40, cmap=cm.Reds, cmin=4,
range = [[0,1],[1,12]])
plt.xlabel(r'$\eta$ (log shootings per year)')
plt.ylabel(r'$\rho^{-1}$ (years)')
sqrt_eta = np.sqrt(stan_model_ext_rho_strong['eta_sq'])
px = np.linspace(min(sqrt_eta), max(sqrt_eta), 10)
px_std = (px - np.mean(sqrt_eta)) / np.std(sqrt_eta)
plt.plot(px,
# Constant term
(est.params[est.model.exog_names.index('const')] +
# Linear term
px * est.params[est.model.exog_names.index('eta')]
* stan_model_ext_rho_strong['inv_rho'].std()) + stan_model_ext_rho_strong['inv_rho'].mean())

plt.axis()
cb = plt.colorbar()
cb.set_label('Number of posterior samples')
plt.title(r'Strong prior on $\rho^{-1}$')

Out[25]:
<matplotlib.text.Text at 0x7f65616fac50>

When we explore the same correlation in the posterior of the model with a weak prior specified on the timescale hyperparameter, we see somewhat different results:

In [26]:
## Assemble data matrices
y = pd.Series(np.log(stan_model_ext_rho_weak['inv_rho'])); y.name = 'inv_rho'
X = pd.DataFrame({
'eta':np.sqrt(stan_model_ext_rho_weak['eta_sq']),
'mu_0':stan_model_ext_rho_weak['mu_0'],
'mu_b':stan_model_ext_rho_weak['mu_b'],
'sigma':np.sqrt(stan_model_ext_rho_weak['sigma_sq']),
'NB_phi_inv':np.sqrt(stan_model_ext_rho_weak['NB_phi_inv']),
})
## Standardize
X = X - X.mean()
X = X / X.std()
y = (y - y.mean()) / y.std()
## Fit linear model using stats models
est = sm.OLS(y, X).fit()
## Print summary
print est.summary2()

plt.figure()
pa = plt.hist2d(np.sqrt(stan_model_ext_rho_weak['eta_sq']),
stan_model_ext_rho_weak['inv_rho'],
bins=40, cmap=cm.Reds, cmin=4,
range = [[0,4],[1,300]])
plt.xlabel(r'$\eta$ (log shootings per year)')
plt.ylabel(r'$\rho^{-1}$ (years)')
sqrt_eta = np.sqrt(stan_model_ext_rho_weak['eta_sq'])
px = np.linspace(min(sqrt_eta), max(sqrt_eta), 10)
px_std = (px - np.mean(sqrt_eta)) / np.std(sqrt_eta)
plt.plot(px,
# Constant term
(est.params[est.model.exog_names.index('const')] +
# Linear term
px * est.params[est.model.exog_names.index('eta')]
* stan_model_ext_rho_weak['inv_rho'].std()) + stan_model_ext_rho_weak['inv_rho'].mean())

plt.axis()
cb = plt.colorbar()
cb.set_label('Number of posterior samples')
plt.title(r'Weak prior on $\rho^{-1}$')

                   Results: Ordinary least squares
====================================================================
Dependent Variable: inv_rho          AIC:                235996.6613
Date:               2017-01-14 23:07 BIC:                236052.6927
No. Observations:   84000            Log-Likelihood:     -1.1799e+05
Df Model:           5                F-statistic:        486.1
Df Residuals:       83994            Prob (F-statistic): 0.00
R-squared:          0.028            Scale:              0.97194
----------------------------------------------------------------------
Coef.    Std.Err.      t      P>|t|     [0.025    0.975]
----------------------------------------------------------------------
const         0.0000     0.0034    0.0000   1.0000   -0.0067    0.0067
NB_phi_inv   -0.0065     0.0034   -1.9194   0.0549   -0.0132    0.0001
eta           0.1635     0.0034   47.9897   0.0000    0.1568    0.1701
mu_0          0.0101     0.0037    2.7591   0.0058    0.0029    0.0172
mu_b         -0.0321     0.0037   -8.7932   0.0000   -0.0393   -0.0250
sigma        -0.0116     0.0034   -3.4000   0.0007   -0.0182   -0.0049
--------------------------------------------------------------------
Omnibus:             4585.854       Durbin-Watson:          1.816
Prob(Omnibus):       0.000          Jarque-Bera (JB):       5770.812
Skew:                -0.547         Prob(JB):               0.000
Kurtosis:            3.673          Condition No.:          1
====================================================================


Out[26]:
<matplotlib.text.Text at 0x7f6561984390>

Again, $\eta$ is the parameter most significantly correlated with $\rho^{-1}$, but now the 2D posterior visualization shows that this correlation is substantially non-linear. In particular for the model with the weak prior on $\rho$, $\eta$ is constrained to much smaller values when the timescale $\rho^{-1}$ is small. In other words, in models that permit variations from the mean function on timescales smaller than the observational range ($\sim35$ years), the amplitude of those variations is constrained to be very small. In any scenario, as we have seen, the importance of the covariance function is minimal under this prior.

## Parameter inferences¶

Below we show the marginalized posterior distributions of the parameters of the Gaussian process under the strong prior on $\rho$.

In [27]:
def gt0(y, x, lbound=0, ubound=np.inf):
y[(x<lbound) & (x>ubound)] = 0
return y

def marg_post_plot(stan_model_ext, alpha_rho, beta_rho, Nhist=25):
hyp_dic = {
'eta_sq': ('$\\eta$', np.sqrt, 'log shootings per year', lambda x: sstats.cauchy.pdf(x**2, 0, 1)),
'inv_rho': ('$\\rho^{-1}$', lambda x: x, 'years', lambda x: gt0(sstats.gamma.pdf(x, alpha_rho, scale=beta_rho), x, lbound=1)),
'sigma_sq': ('$\\sigma$', np.sqrt, 'log shootings per year', lambda x: sstats.cauchy.pdf(x**2, 0, 1)),
'NB_phi_inv':('$\\rm{NB}_\\phi^{-1}$', lambda x:x, '', lambda x: sstats.cauchy.pdf(x**2, 0, 0.5)),
}

meanfunc_dic = {
'mu_0': ('$\\mu_0$', lambda x: x, 'log shootings per year, '+str(np.min(data_years)), lambda x: sstats.norm.pdf(x, 0,2)),
'mu_b': ('$\\mu_b$', lambda x: x, 'annual increase in\nlog shootings per year', lambda x: sstats.norm.pdf(x, 0,0.2)),
}

for name,pdic in (('hyper', hyp_dic), ('meanfunc', meanfunc_dic)):
fig,axs = plt.subplots(1,len(pdic), figsize=(2.5*len(pdic), 2.5), sharey='all')
axs[0].set_ylabel('HMC samples ({} total)'.format(N_samp))
for i,hyp in enumerate(pdic.keys()):
samps = pdic[hyp][1](stan_model_ext[hyp])
hn, hb, hp = axs[i].hist(samps, Nhist, edgecolor='none', facecolor='.5', label='Posterior samples')
ppx = np.linspace(np.min(samps), np.max(samps), 10000)
ppy = pdic[hyp][1]( pdic[hyp][3](ppx) )
## Normalize
ppy *= len(samps) / np.sum(ppy) * len(ppy) / len(hn)
axs[i].plot(ppx, ppy, color='b', zorder=2, label='Hyperprior')
axs[i].xaxis.set_major_locator(MaxNLocator(3))
axs[i].xaxis.set_minor_locator(AutoMinorLocator(3))
axs[i].set_xlabel(pdic[hyp][0] + ' ({})'.format(pdic[hyp][2]), ha='center')
axs[i].axvline(0, ls='dashed', color='.2')
axs[-1].legend(prop={'size':9})

print "Strong prior on rho:"
marg_post_plot(stan_model_ext_rho_strong, stan_data_rho_strong['alpha_rho'], 1/stan_data_rho_strong['beta_rho'], Nhist=100)

Strong prior on rho: