Mr. Wheeler,

Thank you for your page https://andrewpwheeler.com/2016/10/19/testing-the-equality-of-two-regression-coefficients/

I did your technique (at the end of the page) of re-running the model with X+Z and X-Z as independent variables (with coefficients B1 and B2, respectively).

I understand:

- (although you did not say so) that testing whether coefficient b1 (X’s coefficient in the original equation) is LESS THAN coefficient b2 (Z’s coefficient in the original regression) is a one-sided test; and testing whether one coefficient is DIFFERENT from another is a two-sided test
- that the 90%-confidence t-distribution-critical-values-with-infinite-degrees-of-freedom are 1.282 for one-sided tests and 1.645 for two-sided tests
- that if the resulting t-stat for the B2 coefficient is say 1.5, then—according to the tests—I should therefore be 90% confident that b1 is in fact less than b2; and I should NOT be 90% confident that b1 is different from b2.
But—according to MY understanding of logic and statistics—if I am 90% confident that b1 is LESS THAN b2, then I would be MORE THAN 90% confident that b1 DIFFERS from b2 (because “differs” includes the additional chance that b1 is greater than b2), i.e. the tests and my logic conflict. What am I doing wrong?

Rob

So I realize null hypothesis statistical testing (NHST) can be tricky to interpret – but the statement in 3 is not consistent with how we do NHST for several reasons.

So if we have a null hypothesis that Beta1 = Beta2, for reasons to do with the central limit theorem we actually rewrite this to be:

`Null: Beta1 - Beta2 = 0 => Theta0`

I’ve noted this new parameter we are testing – the difference in the two coefficients – as Theta0. For NHST we assume this parameter is 0, and then test to see how close our data is to this parameter. So we estimate with our data:

```
b1 - b2 = Diff
DiffZ = Diff/StandardError_Diff
```

Now, to calculate a p-value, we need to say how unlikely our data estimate, DiffZ, is given the assumed null distribution Theta0. So imagine we draw our standard normal distribution curve about Theta0. This then defines the space for NHST, for a typical two sided test we have (here assuming DiffZ is a negative value):

`P(Z < DiffZ | Theta0 ) + P(Z > -DiffZ | Theta0 ) = Two tailed p-value`

Where less than Z is our partitioning of the space of the null hypothesis, since an exact value for DiffZ here when the distribution of potential outcomes is continuous is zero. For a one sided test, you would just take the relevant portion of the above, and not add the two above portions together:

```
P(Z < DiffZ | Theta0 ) = One tail p-value for Beta1 < Beta2
P(Z > -DiffZ | Theta0 ) = One tail p-value for Beta1 > Beta2
```

Note here that the test is *conditional* on the null hypothesis. Statements such as ‘I should therefore be 90% confident that b1 is in fact less than b2’, which seem to estimate the complement of the p-value (e.g. 1 – p-value) and interpret it as a meaningful probability are incorrect.

P-values are basically numerical summaries of how close the data are to the presumed null distribution. Small p-values just indicate they are not close to the assumed null distribution. The complement of the p-value is not evidence for the alternative hypothesis. It is just the left over distribution for the null hypothesis that is inside the Z values.

Statisticians oftentimes at this point in the conversation suggest Bayesian analysis and instead interpret posteriori probabilities instead of p-values. I will stop here though, as I am not sure “90% confident” readily translates into a specific Bayesian statement. (It could be people are better off doing inferiority/equivalence testing for example, e.g. changing the null hypothesis.)

]]>For a recent example at my work, was working on a model that originally has around 6 million observations and around a dozen categorical inputs (each with fairly high cardinality, e.g. around 1000 different categories). When aggregating to unique cases, this model is well under half a million rows though. It is much easier for me to iterate and fit models on the half a million row dataset than the 6 million row one.

Here I will show an example using NIBRS data. See my prior blog post on Association Rules for background on the data, I will be using the 2012 NIBRS dataset, which has over 5 million observations. Below is python code to illustrate, but pretty much all statistical packages allow you weight observations.

So first I load the libraries I will be using, and make a nice function to print out the logit coefficients for a sklearn model.

```
import pandas as pd
import numpy as np
from scipy.stats import binom
from sklearn.linear_model import LogisticRegression
from sklearn.metrics import roc_auc_score
# Function to pretty print logit coefficients
def nice_coef(mod,x_vars):
coef = list(mod.coef_[0]) + list(mod.intercept_)
vlist = x_vars + ['Intercept']
return pd.DataFrame(zip(vlist,coef),columns=['Var','Coef'])
```

Next we can read in my NIRBS data directly from the dropbox link (replace www with dl to do this in general for dropbox links).

```
# See https://github.com/apwheele/apwheele.github.io/tree/master/MathPosts/association_rules
# For explanation behind NIBRS data
ndum = pd.read_csv('https://dl.dropbox.com/sh/puws33uebzt9ckd/AADVM86qPJVqP4RHWkWfGBzpa/NIBRS_DummyDat.csv?dl=0')
# If you want to read original NIRBS data
# use https://dl.dropbox.com/sh/puws33uebzt9ckd/AACL3wBhZDr3P_ZbsbUxltERa/NIBRS_2012.csv?dl=0
```

This data is already prepped with the repeated dummy variables for different categories. It is over 5 million observations, but a simple way to work with this data is to use a `groupby`

and create a weight variable:

```
group_vars = list(ndum)
ndum['weight'] = 1
ndum_agg = ndum.groupby(group_vars, as_index=False).sum() # sums the weight variable
print(ndum.shape)
print(ndum_agg.shape)
```

So you can see we went from over 5 million observations to only a few over 7000.

A few notes here. One, if querying the data from a DB, it may be better to do the counts on the DB side and only load in the tinier data into memory, e.g. `SELECT COUNT(ID) AS weight, A1,A2... FROM Table GROUP BY A1,A2,....`

.

Second, I do not have any missing data here, but pandas groupby will by default drop missing rows. So you may want to do something like `data.fillna(-1).groupby`

, or the option to not drop NA values.

Now, lets go onto to fitting a model. Here I am using logit regression, as it is easier to compare the inputs for the weighted/non-weighted model, but you can do this for whatever machine learning model you want. I am predicting the probability an officer is assaulted.

```
logit_mod = LogisticRegression(penalty='none', solver='newton-cg')
y_var = 'ass_LEO_Assault'
x_vars = group_vars.copy() #[0:7]
x_vars.remove(y_var)
# Takes a few minutes on my machine!
logit_mod.fit(ndum[x_vars],ndum[y_var])
# Coefficients for the model
bigres = nice_coef(logit_mod,x_vars)
bigres
```

I was not sure if my computer would actually fit this model without running out of memory. But it did crunch it out in a few minutes. Now lets look at the results when we estimate the model with the weighted data. In all the sklearn models you can just pass in a `sample_weight`

into the fit function.

```
logit_mod.fit(ndum_agg[x_vars],ndum_agg[y_var],sample_weight=ndum_agg['weight'])
weight_res = nice_coef(logit_mod,x_vars)
weight_res
```

You can see that these are basically the same model predictions. For a few of the coefficients you can find discrepancies starting at the second decimal, but the majority are within floating point errors.

`bigres['Coef'] - weight_res['Coef']`

This was fit instantly instead of waiting several minutes. For more intense ML models (like random forests), this can dramatically improve the time it takes to fit models.

If you are interested in doing a train/test split, this is quite easy with the weights as well. Basically you just need to allocate some of the weight to the train and some to the test. Here I show how to do that using a binomial random variable. Then you feed the train weights to the fit function:

```
train_prop = 0.5
train_weight = binom.rvs(ndum_agg['weight'].astype(int), train_prop, random_state=10)
test_weight = ndum_agg['weight'] - train_weight
logit_mod.fit(ndum_agg[x_vars],ndum_agg[y_var],sample_weight=train_weight)
```

And in sklearn pretty much all of the evaluation functions also take a sample weight function.

```
pred_probs = logit_mod.predict_proba(ndum_agg[x_vars])
# Eval metrics for the test data
roc_auc_score(ndum_agg[y_var],pred_probs[:,1],sample_weight=test_weight)
roc_auc_score(ndum_agg[y_var],pred_probs[:,1],sample_weight=train_weight)
```

So this shows that the AUC metric decreases in the test dataset (as you would expect it to). Note do not take this model seriously, I would need to think more thoroughly about the selection of rows here, as well as how to effectively interpret these particular categorical inputs in a more reasonable way than eyeballing the coefficients.

I am wondering if weighting the data is actually a more convenient way to do train/test CV splits, just change the weights instead of splitting up datasets. (You could also do fractional weights, e.g. `train_weight = ndum_agg['weight']/2`

, which works nice for stratifying the sample, but may cause some issues generalizing.)

So note this does not always work – but works best with sparse/categorical data. If you have numeric data, you can try to discretize the data to a reasonable amount to still model it as a continuous input (e.g. round age to one decimal, e.g. 20.1). But if you have more than a few numeric inputs you will have a much harder time compressing the data into a smaller number of weighted rows.

It also only works if you have a limited number of inputs. If you have 100 variables you will be able to aggregate less than if you are working with 10.

But despite those limitations, I have come across several different examples in my career where aggregating and weighting the data was clearly the easiest approach, and NIBRS is a great example.

]]>We actually biased our predictions to meet the racial fairness constraint, so you can see we did much better in those categories in Round 1 and Round 2. Unfortunately you only win if you get top in this category – no second place winners here (it says Brier score in these tables, but this is `(1 - BrierScore)*(1 - FPDifference)`

:

But we got lucky and won the overall in Round 2 despite biasing our predictions. Round 3 we have no excuse really, while the predictions were biased it did not matter.

We will do a paper for the results, but overall our approach is pretty standard. For each round we did a grid search over various models – for R1 and R3 we did a L1 logit, for R2 we did an XGBoost model. I did attempt a specialized Logit model with the fairness constraints in the loss function (and just used backpropogation to fit the model, ala deep learning), but in practice the way the fairness metric is done this just added noise into the estimate.

I will have more to say in the future about fairness metrics, unfortunately here I do not think it was well thought out. It was simply the false positive rate comparing white/black subgroups, assuming a threshold of 0.5, which does not make sense in practice. (I’ve written about calculating the threshold for bail here, it applies the same to parole though as well.) So for each model we simply clipped probabilities to be below 0.5 to meet this – no one predicted high means 0 false positives for each group.

So the higher threshold makes it silly, also the multiplication between the metrics I don’t think is a good idea either. I think it can be amended though to be a more reasonable additive fairness constraint. E.g. `BrierScore + lambda*FPDifference`

, where `lambda`

is a tuner to set how you want to make the tradeoff (and FP may be the total N difference, not a proportion difference, which can be volatile for small N). (Also I think it makes more sense to balance false negatives than false positives in the CJ example, but any algorithm to balance one can be flipped to balance the other.)

I do like how NIJ spreads prizes out, instead of Kaggle like with only 1/2/3 big prizes. I wish here we could submit two predictions though (one for main and one for fair). (I am pretty sure we would have placed in Year1 if we did not bias our predictions.)

]]>The chief and mayor wanted a breakdown of particular noteworthy events, so I would place my own notes in a spreadsheet and then make a daily report. My set up was not fully automated but close – I had a pretty detailed HTML template, and once my daily data was inputted, I would run a SPSS script to fill in the HTML. I also did a simple pin map in batch geo (one thing that was not automated about it) and inserted into the report.

I had two other quite regular reports I would work on. One was a weekly command staff report about overall trends and recent upticks, the other was a monthly Compstat meeting going over similar stats. (I also had various other products to release – detective assignments/workload, sending aggregate stats to the Albany Crime Analysis Center.)

If I had to do these again knowing what I know now, I would automate nearly 100% of this work in python. For the reports, I would use jupyter notebooks (I actually do not like coding in these very much, I much prefer plain text IDEs, but they are good for generating nice looking reports I will show.)

I have provided the notes to fully automate a simple report here on Github. To replicate, first you need to download the Dallas PD open data and create a local sqlite database (can’t upload that large of file to github).

So first before you start, if you download the `.py`

files, you can run at the command prompt something like:

```
cd D:\Dropbox\Dropbox\PublicCode_Git\Blog_Code\Python\jupyter_reports
python 00_CreateDB.py
```

Just replace the `cd`

path to wherever you saved the files on your local machine (and this assumes you have Anaconda installed). Then once that is done, you can replicate the report locally, but it is really meant as a pedogological tool – you can see how I wrote the jupyter notebook to query the local database. For your own case you would write the SQL code and connection to whereever your local crime data is store.

Here is an example of how you can use string substitution in python to create a query that is date aware for when the code is run:

Part of the hardest part of making standardized reports is you can typically make the data formatted how you want, but to get little pieces looking exactly how you want them is hard. So here is a default pivot table exported in a Jupyter notebook for some year to date statistics (note the limitations of this, why I prefer graphs and do not show a percent change).

And here is code I wrote to change font sizes and insert a title. A bit of work, but once you figure it out once you are golden.

You can go look at the notebook itself, but I also have an example of generating a weekly error bar chart (much preferred over the Compstat YTD tables):

Final note, to compile the notebook without showing any code (the police chief does not want to see your python code!), it looks like this from the command line.

`jupyter nbconvert --execute example_report.ipynb --no-input --to html`

I have further notes in the github page on automating this fully via bat files for windows, renaming files to make them update to the current date, etc.

I know some analysts are reading this and thinking to themselves – I can generate these reports in 30 minutes using Excel and Powerpoint – why spend time to learn something new to make it 100% automated? There are a few reasons. One is pure time savings in the end. Say for the weekly report you spend 1 hour, and it takes you three work days (24 hours) to fully automate. You will recover your time in 24 weeks.

Time savings is not the only component though. Fully automating means the workflow is 100% reproducible, and makes it easier to transfer that workflow to other analysts in the future. This is an important consideration when *scaling* – if you need to spend a few hours once a week forever, you can only take on generating so many reports. It is better for your time to do a one time large sink into automating something, than to loan out a portion of your time forever.

A final part is the skills you develop when generating automated reports are more similar to data science roles in the private sector – so consider it an investment in your career as well. The types of machine learning pipelines I create at my current role would not be possible if I could not fully automate. I would only be able to do 2 or maybe 3 projects forever and just maintain them. I fully automate my data pipelines though, and then can hand off that job to a DevOps engineer, and only worry about fixing things when they break. (Or a more junior data scientist can take over that project entirely.)

]]>As a follow up to my prior post on spatial sample size recommendations for the SPPT test, I figured I would show an actual analysis of spatial changes in crime. I’ve previously written about how NYC shootings appear to be going up by a similar amount in each precinct. We can do a similar analysis, but at smaller geographic spatial units, to see if that holds true for everywhere.

The data and R code to follow along can be downloaded here. But I will copy-paste below to walk you through.

So first I load in the libraries I will be using and set my working directory:

```
###################################################
library(sppt)
library(sp)
library(raster)
library(rgdal)
library(rgeos)
my_dir <- 'C:\\Users\\andre\\OneDrive\\Desktop\\NYC_Shootings_SPPT'
setwd(my_dir)
###################################################
```

Now we just need to do alittle data prep for the NYC data. Concat the old and new files, convert the data fields for some of the info, and do some date manipulation. I choose the pre/post date here March 1st 2020, but also note we had the Floyd protests not to long after (so calling these Covid vs protest increases is pretty much confounded).

```
###################################################
# Read in the shooting data
old_shoot <- read.csv('NYPD_Shooting_Incident_Data__Historic_.csv', stringsAsFactors=FALSE)
new_shoot <- read.csv('NYPD_Shooting_Incident_Data__Year_To_Date_.csv', stringsAsFactors=FALSE)
# Just one column off
print( cbind(names(old_shoot), names(new_shoot)) )
names(new_shoot) <- names(old_shoot)
shooting <- rbind(old_shoot,new_shoot)
# I need to conver the coordinates to numeric fields
# and the dates to a date field
coord_fields <- c('X_COORD_CD','Y_COORD_CD')
for (c in coord_fields){
shooting[,c] <- as.numeric(gsub(",","",shooting[,c])) #replacing commas in 2018 data
}
# How many per year to check no funny business
table(substring(shooting$OCCUR_DATE,7,10))
# Making a datetime variable in R
shooting$OCCUR_DATE <- as.Date(shooting$OCCUR_DATE, format = "%m/%d/%Y", tz = "America/New_York")
# Making a post date to split after Covid started
begin_date <- as.Date('03/01/2020', format="%m/%d/%Y")
shooting$Pre <- ifelse(shooting$OCCUR_DATE < begin_date,1,0)
#There is no missing data
summary(shooting)
###################################################
```

Next I read in a shapefile of the census tracts for NYC. (Pro-tip for NYC GIS data, I like to use Bytes of the Big Apple where available.) The interior has a few dongles (probably for here should have started with a borough outline file), so I do a tiny buffer to get rid of those interior dongles, and then smooth the polygon slightly. To check and make sure my crime data lines up, I superimpose with a tiny dot map — this is also a great/simple way to see the overall shooting density without the hassle of other types of hot spot maps.

```
###################################################
# Read in the census tract data
nyc_ct <- readOGR(dsn="nyct2010.shp", layer="nyct2010")
summary(nyc_ct)
plot(nyc_ct)
nrow(nyc_ct) #2165 tracts
# Dissolve to a citywide file
nyc_ct$const <- 1
nyc_outline <- gUnaryUnion(nyc_ct, id = nyc_ct$const)
plot(nyc_outline)
# Area in square feet
total_area <- area(nyc_outline)
# 8423930027
# Turning crimes into spatial point data frame
coordinates(shooting) <- coord_fields
crs(shooting) <- crs(nyc_ct)
# This gets rid of a few dongles in the interior
nyc_buff <- gBuffer(nyc_outline,1,byid=FALSE)
nyc_simpler <- gSimplify(nyc_buff, 500, topologyPreserve=FALSE)
# Checking to make sure everything lines up
png('NYC_Shootings.png',units='in',res=1000,height=6,width=6,type='cairo')
plot(nyc_simpler)
points(coordinates(shooting),pch='.')
dev.off()
###################################################
```

The next part I created a function to generate a nice grid over an outline area of your choice to do the SPPT analysis. What this does is generates the regular grid, turns it from a raster to a vector polygon format, and then filters out polygons with 0 overlapping crimes (so in the subsequent SPPT test these areas will all be 0% vs 0%, so not much point in checking them for differences over time!).

You can see the logic from the prior blog post, if I want to use the area with power to detect big changes, I want `N*0.85`

. Since I am comparing data over 10 years compared to 1+ years, they are big differences, so I treat N here as 1.5 times the newer dataset, which ends up being around a suggested 3,141 spatial units. Given the area for the overall NYC, this translates to grid cells that are about 1600 by 1600 feet. Once I select out all the 0 grid cells, there only ends up being a total of 1,655 grid cells for the final SPPT analysis.

```
###################################################
# Function to create sppt grid over areas with
# Observed crimes
grid_crimes <- function(outline,crimes,size){
# First creating a raster given the outline extent
base_raster <- raster(ext = extent(outline), res=size)
projection(base_raster) <- crs(outline)
# Getting the coverage for a grid cell over the city area
mask_raster <- rasterize(outline, base_raster, getCover=TRUE)
# Turning into a polygon
base_poly <- rasterToPolygons(base_raster,dissolve=FALSE)
xy_df <- as.data.frame(base_raster,long=T,xy=T)
base_poly$x <- xy_df$x
base_poly$y <- xy_df$y
base_poly$poly_id <- 1:nrow(base_poly)
# May also want to select based on layer value
# sel_poly <- base_poly[base_poly$layer > 0.05,]
# means the grid cell has more than 5% in the outline area
# Selecting only grid cells with an observed crime
ov_crime <- over(crimes,base_poly)
any_crime <- unique(ov_crime$poly_id)
sub_poly <- base_poly[base_poly$poly_id %in% any_crime,]
# Redo the id
sub_poly$poly_id <- 1:nrow(sub_poly)
return(sub_poly)
}
# Calculating suggested sample size
total_counts <- as.data.frame(table(shooting$Pre))
print(total_counts)
# Lets go with the pre-total times 1.5
total_n <- total_counts$Freq[1]*1.5
# Figure out the total number of grid cells
# Given the total area
side <- sqrt( total_area/total_n )
print(side)
# 1637, lets just round down to 1600
poly_cells <- grid_crimes(nyc_simpler,shooting,1600)
print(nrow(poly_cells)) #1655
png('NYC_GridCells.png',units='in',res=1000,height=6,width=6,type='cairo')
plot(nyc_simpler)
plot(poly_cells,add=TRUE,border='blue')
dev.off()
###################################################
```

Next part is to split the data into pre/post, and do the SPPT analysis. Here I use all the defaults, the Chi-square test for proportional differences, along with a correction for multiple comparisons. Without the multiple comparison correction, we have a total of 174 grid cells that have a p-value < 0.05 for the differences in proportions for an S index of around 89%. With the multiple comparison correction though, the majority of those p-values are adjusted to be above 0.05, and only 25 remain afterwards (98% S-index). You can see in the screenshot that *all* of those significant differences are increases in proportions from the pre to post. While a few are 0 shootings to a handful of shootings (suggesting diffusion), the majority are areas that had multiple shootings in the historical data, they are just at a higher intensity now.

```
###################################################
# Now lets do the sppt analysis
split_shoot <- split(shooting,shooting$Pre)
pre <- split_shoot$`1`
post <- split_shoot$`0`
library(dplyr)
sppt_diff <- sppt_diff(pre, post, poly_cells)
summary(sppt_diff)
# Unadjusted vs adjusted p-values
sum(sppt_diff$p.value < 0.05) #174, around 89% similarity
sum(sppt_diff$p.adjusted < 0.05) #25, 98% similarity
# Lets select out the increases/decreases
# And just map those
sig <- sppt_diff$p.adjusted < 0.05
sppt_sig <- sppt_diff[sig,]
head(sppt_sig,25) # to check out all increases
###################################################
```

The table is not all that helpful though for really digging into patterns, we need to map out the differences. The first here is a map showing the significant grid cells. They are somewhat tiny though, so you have to kind of look close to see where they are. The second map uses proportional circles to the percent difference (so bigger circles show larger increases). I am too lazy to do a legend/scale, but see my prior post on a hexbin map, or the sp website in the comments.

```
###################################################
# Making a map
png('NYC_SigCells.png',units='in',res=1000,height=6,width=6,type='cairo')
plot(nyc_simpler,lwd=1.5)
plot(sppt_sig,add=TRUE,col='red',border='white')
dev.off()
circ_sizes <- sqrt(-sppt_sig$diff_perc)*3
png('NYC_SigCircles.png',units='in',res=1000,height=6,width=6,type='cairo')
plot(nyc_simpler,lwd=1.5)
points(coordinates(sppt_sig),pch=21,cex=circ_sizes,bg='red')
dev.off()
# check out https://edzer.github.io/sp/
# For nicer maps/legends/etc.
###################################################
```

So the increases appear pretty spread out. We have a few notable ones that made the news right in the thick of things in Manhattan, but there are examples of grid cells that increased scattered all over the boroughs. I am not going to the trouble here, but if I were a crime analyst working on this, I would export this to a format where I could zoom into the local areas and drill down into the specific incidents. You can do that either in ArcGIS, or more directly in R by creating a leaflet map.

So if folks have any better ideas for testing out crime increases I am all ears. At some point will give the R package sparr a try. (Here you could treat pre as the controls and post as the cases.) I am not a real big fan of over interpreting changes in kernel density estimates though (they can be quite noisy, and heavily influenced by the bandwidth), so I do like the SPPT analysis by default (but it swaps out a different problem with choosing a reasonable grid cell size).

]]>Long story short, if you have N crimes, I think you should either use `0.85*N = S`

spatial units of analysis at the high end, but can only detect very large changes. To be well powered to detect smaller changes between the two distributions, use `0.45*N = S`

. That is, if you have two crime samples you want to compare, and the smaller sample has 1000 crimes, the largest spatial sample size I would recommend is 850 units, but I think 450 units is better.

For those not familiar with the SPPT technique, it compares the proportion of events falling inside a common area (e.g. police beats, census block groups, etc.) between two patterns. So for example in my work I compared the proportion of violent crime and the proportion of SQF in New York City (Wheeler et al., 2018). I think it makes sense as a gross monitoring metric for PDs this way (say for those doing DDACTS, swap out pedestrian stops with traffic stops), so you can say things like `area A had a much lower proportion of crimes than stops, so we should emphasize people do fewer stops in A overall`

.

If you are a PD, you may already know the spatial units you want to use for monitoring purposes (say for each police sector or precinct). In that case, you want the power analysis to help guide you for how large a sample you need to effectively know how often you can update the estimates (e.g. you may only have enough traffic stops and violent crimes to do the estimates on a quarterly basis, not a monthly one) Many academic papers though are just generally theory testing, so don’t have an a priori spatial unit of analysis chosen. (But they do have two samples, e.g. a historical sample of 2000 shootings and a current sample of 1000 shootings.) See Martin’s site for a list of prior papers using the SPPT to see it in action.

I’ve reviewed several papers that examine these proportion changes using the SPPT at very tiny spatial units of analysis, such as street segments. They also happen to have very tiny numbers of overall crimes, and then break the crimes into subsequent subsets. For example reviewed a paper that had around 100 crimes in each subset of interest, and had around 20,000 street segments. I totally get wanting to examine micro place crime patterns – but the SPPT is not well suited for this I am afraid.

Ultimately if you chunk up the total number of crimes into smaller and smaller areas, you will have less statistical power to uncover differences. With very tiny total crime counts, you will be basically only identifying differences between areas that go from 0 to 1 or 0 to 2 etc. It also becomes much more important to control for multiple comparisons when using a large number of spatial units. In general this technique is not going to work out well for micro units of analysis, it will only really work out for larger spatial units IMO. But here I will give my best advice about how small you can reasonably go for the analysis.

There are quite a few different ways people have suggested to determine the spatial sample for areas when conducting quadrat analysis (e.g. when you make your own spatial areas). So one rule of thumb is to use `2*A/N`

, where A is the area of the study and N is the total number of events (Paez, 2021).

Using the SPPT test itself, Malleson et al. (2019) identify the area at which the spatial pattern exhibits the highest similarity index with itself using a resampling approach. Ramos et al. (2021) look at the smallest spatial unit at which the crime patterns within that unit show spatial randomness.

So those later two take an error metric based approach (the spatial unit of analysis likely to result in the miminal amount of error, with error defined different ways). I take a different approach here – power analysis. We want to compare to spatial point patterns for proportional differences, how can we construct the test to be reasonably powered to identify differences we want to detect?

I do not have a perfect way to do this power analysis, but here is my logic. Crime patterns are often slightly overdispersed, so here I assume if you split up say 1000 crimes into 600 areas, it will have an NB2 distribution with a mean of `1000/600 = 1.67`

and an overdispersion parameter of 2. (I assume this parameter to be 2 for various reasons, based on prior analysis of crime patterns, and that 2 tends to be in the general ballpark for the amount of overdispersion.) So now we want to see what it would take to go from a hot spot of crime, say the 98th percentile of this distribution to the median 50th percentile.

So in R code, to translate the NB2 mean/dispersion to N & P notation results in N & P parameters of `1.0416667`

and `0.3846154`

respectively:

```
trans_np <- function(mu,disp){
a <- disp
x <- mu^2/(1 - mu + a*mu)
p <- x/(x + mu)
n <- (mu*p)/(1-p)
return(c(n,p))
}
# Mean 1000/600 and dispersion of 2
nb_dis <- trans_np(1000/600,2)
```

Now we want to see what the *counts* are to go from the 98th to the 50th percentile of this distribution:

`crime_counts <- qnbinom(c(0.98,0.5), size=nb_dis[1], prob=nb_dis[2])`

And this gives us a result of `[1] 8 1`

in the `crime_counts`

object. So a hot spot place in this scenario will have around 8 crimes, and the median will be around 1 crime in our hypothetical areas. So we can translate these to percentages, and then feed them into R’s `power.prop.test`

function:

```
crime_prop <- crime_counts/1000
power.prop.test(n = 1000, p1 = crime_prop[1], p2 = crime_prop[2])
```

And this gives us a result of:

```
Two-sample comparison of proportions power calculation
n = 1000
p1 = 0.008
p2 = 0.001
sig.level = 0.05
power = 0.6477139
alternative = two.sided
NOTE: n is number in *each* group
```

Note that this is for one N estimate, and assumes that N will be the same *for each* proportion. In practice for the SPPT test this is not true, oftentimes we have two crime samples (or crime vs police actions like stops), which have very different total baseline N’s. (It is part of the reason the test is useful, it doesn’t make so much sense in that case to compare densities as it does proportions.) So subsequently when we do these estimates, we should either take the *average* of the total number of crimes we have in our two point patterns for SPPT (if they are close to the same size), or the *minimum* number of events if they are very disparate. So if you have in sample A 1000 crimes, and sample B 2000 crimes, I think you should treat the N in this scenario as 1500. If you have 5000 crimes vs 1000000 crimes, you should treat N here as 5000.

So that estimate above is for one set of crimes (1000), and one set of areas (600). But what if we vary the number of areas? At what number of areas do we have the maximum power?

So I provide functions below to generate the power estimate curve, given these assumptions about the underlying crime distribution (which will generally be in the ballpark for many crime patterns, but not perfect), for varying numbers of spatial units. Typically we know the total number of crimes, so we are saying given I have N crimes, how finely can a split them up to check for differences with the SPPT test.

Both the Malleson and Ramos article place their recommendations in terms of area instead of total number of units. But it would not surprise me if our different procedures end up resulting in similar recommendations based on the observed outputs of each of the papers. (The `2A/N`

quadrat analysis suggestion translates to `N*0.5`

total number of areas, pretty close to my `0.45*N`

suggestion for example.)

Below I have a nicer function to do the analysis I walked through above, but give a nice power curve and dataframe over various potential spatial sample sizes:

```
# SPPT Power analysis example
library(ggplot)
# See https://andrewpwheeler.com/2015/01/03/translating-between-the-dispersion-term-in-a-negative-binomial-regression-and-random-variables-in-spss/
trans_np <- function(mu,disp){
a <- disp
x <- mu^2/(1 - mu + a*mu)
p <- x/(x + mu)
n <- (mu*p)/(1-p)
return(c(n,p))
}
diff_suggest <- function(total_crimes,areas=round(seq(2,total_crimes*10,length.out=500)),
change_quant=c(0.98,0.5), nb_disp=2, alpha = 0.05,
plot=TRUE){
# Figure out mean
areas <- unique(areas)
mean_cr <- total_crimes/areas
# Initialize some vectors to place the results
n_areas <- length(areas)
power <- vector("numeric",length=n_areas)
hign <- power
lown <- power
higp <- power
lowp <- power
# loop over areas and calculate power
for (i in 1:length(areas)){
# Negative binomial parameters
dp <- trans_np(mean_cr[i],nb_disp)
hilo <- qnbinom(change_quant, size=dp[1], prob=dp[2])
hilo_prop <- hilo/total_crimes
# Power for test
pow <- power.prop.test(n = total_crimes, p1 = hilo_prop[1], p2 = hilo_prop[2],
sig.level = alpha)
# Stuffing results in vector
power[i] <- pow$power
hign[i] <- hilo[1]
lown[i] <- hilo[2]
higp[i] <- hilo_prop[1]
lowp[i] <- hilo_prop[2]
}
Ncrimes <- rep(total_crimes,n_areas)
res_df <- data.frame(Ncrimes,areas,mean_cr,power,hign,lown,higp,lowp)
# replacing missing with 0
res_df[is.na(res_df)] <- 0
if (plot) {
require(ggplot2)
fmt_cr <- formatC(total_crimes, format="f", big.mark=",", digits=0)
title_str <- paste0("Power per area for Total number of crimes: ",fmt_cr)
cap_str <- paste0("NB Dispersion = ",nb_disp,", alpha = ",alpha,
", change quantiles = ",change_quant[1]," to ",change_quant[2])
p <- ggplot(data=res_df,aes(x=areas,y=power)) + geom_line(size=1.5) +
theme_bw() + theme(panel.grid.major = element_line(linetype="dashed")) +
labs(x='Number of Areas',y=NULL,title=title_str,caption=cap_str) +
scale_x_continuous(minor_breaks=NULL) + scale_y_continuous(minor_breaks=NULL) +
theme(text = element_text(size=16), axis.title.y=element_text(margin=margin(0,10,0,0)))
print(p)
}
return(res_df)
}
```

Once that function is defined, you can make a simple call like below, and it gives you a nice graph of the power given different numbers of grid cells:

`diff_suggest(100)`

So you can see here we never have very high power over this set of parameters. It is also non-monotonic and volatile at very small numbers of spatial units of analysis (at which the overdispersion assumption likely does not hold, and probably is not of much interest). But once that volatility tamps down we have stepped curve, that ends up happening to step whenever the original NB distribution changes from particular integer values.

So what happens with the power curve if we up the number of crimes to 3000?

`diff_suggest(3000)`

So those two patterns are quite similar. It happens that when breaking down to the smaller units, the highest power scenario is when the crimes are subdivided into around 0.85 *fewer* spatial units than total crimes. So if you have 1000 crimes, in this scenario I would suggest to use 850 areas.

Also note the behavior when you break it down into a very large number of spatial units S, where `S >> N`

, you get a progressive decline until around 0 power in this analysis. E.g. if you have 100 times more spatial units than observations, only a handful of locations have any crimes, and the rest are all 0’s. So you need to be able to tell the difference between 1/N and 0, which is tough (and any inferences you do make will just pretty much be indistinguishable from noise).

What about if we change the quantiles we are examining, and instead of looking at the very high crime place to the median, look at the 80th percentile to the 20th percentile:

`diff_suggest(3000, change_quant=c(0.8,0.2))`

We have a similar step pattern, but here the power is never as high as before, and only maxes out slightly above 0.5. It happens that this 0.5 power is around `number of crimes*0.45`

. So this suggests that to uncover more middling transitions, one would need to have less than half the number of spatial units of crime observed. E.g. if you have 1000 crimes, I would not suggest any more than 450 spatial units of analysis.

So the first scenario, `crimes*0.85`

you could say something like this is the highest power scenario to detect changes from very high crime locations (aka hot spots), to the middle of the distribution. For the second scenario (my preferred offhand), is to say `crimes*0.45`

total spatial units results in the highest power scenario to detect more mild changes in the middle of the distribution (and detecting changes from hot spots to cold spots thus have even more power).

For now that is the best advice I can give for determining the spatial sample size for the SPPT test. Will have a follow up blog post on using R to make a grid to conduct the test.

Also I have been wondering about the best way to quantify changes in the overall ranking. I have not come upon a great solution I am happy with though, so will need to think about it some more.

- Andresen, M. A. (2016). An area-based nonparametric spatial point pattern test: The test, its applications, and the future.
*Methodological Innovations*, 9, 2059799116630659. - Malleson, N., Steenbeek, W., & Andresen, M. A. (2019). Identifying the appropriate spatial resolution for the analysis of crime patterns.
*PloS One*, 14(6), e0218324. - Paez, A. (2021).
*Applied Spatial Statistics with R*. - Ramos, R. G., Silva, B. F., Clarke, K. C., & Prates, M. (2021). Too fine to be good? Issues of granularity, uniformity and error in spatial crime analysis.
*Journal of Quantitative Criminology*, 37(2), 419-443. - Wheeler, A. P., Steenbeek, W., & Andresen, M. A. (2018). Testing for similarity in area-based spatial patterns: Alternative methods to Andresen’s spatial point pattern test. Transactions in GIS, 22(3), 760-774.

The paper examines the increase in case clearances (almost always arrests in this sample) for incidents that occurred nearby 329 public CCTV cameras installed and monitored by the Dallas PD from 2014-2017. Quite a bit of the criminological research on CCTV cameras has examined crime reductions after CCTV installations, which the outcome of that is a consistent small decrease in crimes. Cameras are often argued to help solve cases though, e.g. catch the guy in the act. So we examined that in the Dallas data.

We did find evidence that CCTV increases case clearances on average, here is the graph showing the estimated clearances *before* the cameras were installed (based on the distance between the crime location and the camera), and the line after. You can see the bump up for the post period, around 2% in this graph and tapering off to an estimate of no differences before 1000 feet.

When we break this down by different crimes though, we find that the increase in clearances is mostly limited to theft cases. Also we estimate counterfactual how many extra clearances the cameras were likely to cause. So based on our model, we can say something like, a case would have an estimated probability of clearance without a camera of 10%, but with a camera of 12%. We can then do that counterfactual for many of the events around cameras, e.g.:

```
Probability No Camera Probability Camera Difference
0.10 0.12 + 0.02
0.05 0.06 + 0.01
0.04 0.10 + 0.06
```

And in this example for the three events, we calculate the cameras increased the *total* expected number of clearances to be `0.02 + 0.01 + 0.06 = 0.09`

. This marginal benefit changes for crimes mostly depends on the distance to the camera, but can also change based on when the crime was reported and some other covariates.

We do this exercise for all thefts nearby cameras post installation (over 15,000 in the Dallas data), and then get this estimate of the cumulative number of extra theft clearances we attribute to CCTV:

So even with 329 cameras and over a year post data, we only estimate cameras resulted in fewer than 300 additional *theft* clearances. So there is unlikely any reasonable cost-benefit analysis that would suggest cameras are worthwhile for their benefit in clearing additional cases in Dallas.

For those without access to journals, we have the pre-print posted here. The analysis was not edited any from pre-print to published, just some front end and discussion sections were lightly edited over the drafts. Not sure why, but this pre-print is likely my most downloaded paper (over 4k downloads at this point) – even in the good journals when I publish a paper I typically do not get 1000 downloads.

To go on, complaint number 5631 about peer review – this took quite a while to publish because it was rejected on R&R from Justice Quarterly, and with me and Yeondae both having outside of academia jobs it took us a while to do revisions and resubmit. I am not sure the overall prevalence of rejects on R&R’s, I have quite a few of them though in my career (4 that I can remember). The dreaded send to new reviewers is pretty much guaranteed to result in a reject (pretty much asking to roll a Yahtzee to get it past so many people).

We then submitted to a lower journal, The American Journal of Criminal Justice, where we had reviewers who are not familiar with what counterfactuals are. (An irony of trying to go to a lower journal for an easier time, they tend to have much worse reviewers, so can sometimes be not easier at all.) I picked it up again a few months ago, and re-reading it thought it was too good to drop, and resubmitted to the Journal of Experimental Criminology, where the reviews were reasonable and quick, and Wesley Jennings made fast decisions as well.

]]>What is the best method to examine whether there are group differences (e.g., gender, race) in the effects of several variables on binary outcomes (using logistic regression)? For example – if you want to look at the gendered effects of different types of trauma experiences on subsequent adverse behaviors (e.g., whether participant uses drugs, alcohol, has mental health diagnosis, has attempted suicide). Allison (1999) cautions against using Equality of Coefficients tests to look at group differences between regression coefficients like we might with OLS regression. If you wanted to look at the differences between a lot of predictors (n= 16) on various outcomes (n=6) – what would be best way to go about it (I know using interaction terms would be good if you were just interested in say gender differences of one or two variables on the outcome). Someone recommended comparing marginal effects through average discrete changes (ADCs) or discrete changes at representative values (DCRs) – which is new to me. Would you agree with this suggestion?

When I am thinking about should I use method X or method Y type problems, I think about the specific value I want to estimate first, and then work backwards about the best method to use to get that estimate. So if we are talking about binary endpoints such as uses drugs (will go with binary for now, but my examples will readily extend to say counts or real valued outcomes), there are only generally two values people are interested in; say the change in probability is 5% given some input (absolute risk change, e.g. 10% to 5%), or a relative risk change such as X decreases the overall relative risk by 20% (e.g. 5% to 4%).

The former, absolute change in probabilities, is best accomplished via various marginal effects or average discrete changes as Calli says. Most CJ examples I am aware of I think these make the most sense to focus on, as they translate more directly to costs and benefits. Focusing on the ratio’s sometimes makes sense, such as in case-control studies, or if you want to extrapolate coefficient estimates to a very different sample. Or if you are hyper focused on theory and the underlying statistical model.

Will show an example in Stata using simulated data to illustrate the differences. Stata is very nice to work with different types of marginal estimates. Here I generate data with three covariates, female/males, and then some interactions. Note the covariate x1 has the same effect for males/females, x2 and x3 though have countervailing effects (females it decreases, males it increases the probability).

```
* Stata simulation to show differences in Wald vs Margins
clear
set more off
set seed 10
set obs 5000
generate female = rbinomial(1,0.5)
generate x1 = rnormal(0,1)
generate x2 = rnormal(0,1)
generate x3 = rnormal(0,1)
* x1 same effect, x2/x3 different across groups
generate logit = -0.1 + -2.8*female + 1.1*(x1 + x2 + x3) + -1.5*female*(x2 + x3)
generate prob = 1/(1 + exp(-1*logit))
generate y = rbinomial(1,prob)
drop logit prob
```

I intentionally generated the groups so females/males have quite different baseline probabilities for the outcome y here – something that happens in real victim data in criminology.

```
* Check out marginal differences
tabstat y, by(female)
```

So you can see males have the proportion of the outcome near 50% in the sample, whereas females are under 10%. So Calli is basically interested in the case below, where we estimate all pairwise interactions, so have many coefficient differences to test on the right hand side.

```
* Estimate model with interactions (linear coefficients)
logistic y i.female x1 x2 x3 i.female#(c.x1 c.x2 c.x3), coef
```

This particular model does the Wald tests for the coefficient differences just by the way we have set up the model. So the interaction effects test for differences from the baseline model, so can see the interaction for x1 is not stat significant, but x2/x3 are (as they should be). But if you are interested in the marginal effects, one place to start is with derivatives directly, and Stata automatically for logit models spits out probabilities:

```
* Marginal effects
margins female, dydx(x1 x2 x3)
* x1 is the same linear effect, but margins are quite different
```

So even though I made the underlying effect for x1 equal between males/females for the true underlying data generation process, you can see here the marginal derivative is much smaller for females. This is the main difference between Wald tests and margins.

This is ok though for many situations. Say x1 is a real valued treatment, such as Y is victimization in a sample of high risk youth, and x1 is hours given for a summer job. We want to know the returns of expanding the program – here the returns are higher for males than females due to the different baseline probabilities of risk between the two. This is true even if the relative effect of summer job hours is the same between the two groups.

Again Stata is very convenient, and we can test for the probability differences in males/females by appending `r.`

to the front of the margins command.

```
* can test difference contrast in groups
margins r.female, dydx(x1 x2 x3)
```

But the marginal derivative can be difficult to interpret – it is the average slope, but what does that mean exactly? So I like evaluating at fixed points of the continuous variable. Going back to our summer job hours example, you could evaluate going from 0 to 50 hours, or going from 50 to 100, or 0 to 100, etc. and see the average returns in terms of reductions in the probability of trauma. Here because I simulated the data to be a standard normal, I just go from -1 to 0 to 1:

```
* Probably easier to understand at particular x1 values
margins female, at(x1=(-1(1)1))
```

So that table is dense, but it says when x1=-1, females have a probability of y of 2%, and males have a probability of y of 32%. Now go up the ladder to x1=0, females have a probability of 6% and males have a probability of 48%. So a discrete change of 4% for females and 16% for males. If we want to generate an interval around that discrete change effect:

```
* Can test increases one by one
margins female, at(x1=(-1 0 1)) contrast(atcontrast(ar) effects marginswithin)
```

See, isn’t Stata’s margins command so nice! (For above, it actually may make more sense to use `margins , at(x1=(-1 0 1)) over(female) contrast(atcontrast(ar) effects)`

. Margins estimates the changes over the whole sample and averages filling in certain values, with over it only does the averaging within each group on over.) And finally we can test the difference in difference, to see if the discrete changes in males females from going from -1 to 0 to 1 are themselves significant:

```
* And can test increases of males vs females
margins r.female, at(x1=(-1(1)1)) contrast(atcontrast(ar))
```

So the increase in females is 13% points smaller than the increase in males going from -1 to 0, etc.

So I have spent alot of time on the probabilities so far. I find them much easier to interpret, and I do not care so much about the fact that it doesn’t necessarily say the underlying DGP is different from males/females. But many people are interested in the odds ratios (say in case-control studies). Or generalizing to a different sample, say this is a low risk sample of females, and you want to generalize the odds ratio’s to a higher risk sample with a baseline more around 50%. Then looking at the odds ratio may make more sense.

Or so far I have not even gotten to Calli’s main point, how to test many of these effects for no differences at once. There I would just suggest the likelihood ratio test (which does not have the problems with the Wald tests on the coefficients and that the variance estimates may be off):

```
* Estimate restricted model
logistic y ibn.female c.x1 c.x2 c.x3, coef noconstant
estimates store restrict
* Another way to do the full interaction model
* More like separate male and female
logistic y ibn.female ibn.female#(c.x1 c.x2 c.x3), coef noconstant
estimates store full
* LRT test between models
lrtest restrict full
```

So here as expected, one rejects the null that the restricted model is a better fit to the data. But this is pretty uninformative – I rather just go to the more general model and quantify the differences.

So if you *really* want to look at the odds ratios, we can do that using `lincom`

post our full interaction model:

`logistic y i.female ibn.female#(c.x1 c.x2 c.x3), coef noconstant`

And here is an example post Wald test for equality:

`lincom 0.female#c.x1 - 1.female#c.x1`

You may ask where does this odd’s ratio of 0.921 come from? Well, way back in our first full model, the interaction term for `female*x1`

is `0.0821688`

, and `exp(-0.0821688)`

equals that odds ratio and has the same p-value as the original model I showed. And so you can see that the x1 effect is the same across each group. But estimating the other contrasts is not:

`lincom 0.female#c.x2 - 1.female#c.x2`

And Stata defaults this to estimating a difference in the odds ratio (as far as I can tell you *can’t* tell Stata to do the linear coefficient after logit, would need to change the model to `glm y x, family(binomial) link(logit)`

to do the linear tests).

I honestly don’t know how to really interpret this – but I have been asked for it several different times by clients. Maybe they know better than me, but I think it is more to do with people just expect to be dealing with odds ratios after a logistic regression.

So we can coerce margins to give us odds ratios:

```
* For the odds ratios
quietly margins female, at(x3=(-1(0.1)1)) expression(exp(predict(xb)))
marginsplot , yscale(log) ylabel(0.125 0.25 0.5 1 2 4 8)
```

Or give us the differences in the odds ratio:

```
* For the contrast in the OR
quietly margins r.female, at(x3=(-1(0.1)1)) expression(exp(predict(xb)))
marginsplot
```

(Since it is a negative number cannot be drawn on a log scale.) But again I find it much easier to wrap my head around probabilities:

```
* For the probabilities
quietly margins female, at(x3=(-2(0.1)2))
marginsplot
```

So here while x3 increases, for males it increases the probability and females it decreases. The female decrease is smaller due to the smaller baseline risk in females.

So while Calli’s original question was how to do this reasonably for many different contrasts, I would prefer the actual empirical estimates of the differences. Doing a single contrast among a small number of representative values over many variables and placing in a table/graph I think is the best way to reduce the complexity.

I just don’t find the likelihood ratio tests for all equalities that informative, and for large samples they will almost always say the more flexible model is better than the restricted model.

We estimate models to actually look at the quantitative values of those estimates, not to do rote hypothesis testing.

]]>It’s fascinating to be doing completely unfundable research in the modern university. It means you don’t matter to administration. At all. You are completely irrelevant. You add no value. This means almost all humanities people and a good number of social scientists, though by no means all. Because universities want those corporate dollars, you are encouraged to do whatever corporations want. Bring in that money. But why would we trust any research funded by corporate dollars? The profit motive makes the research inherently questionable. Like with the racism inherent in science and technology, all researchers bring their life experiences into their research. There is no “pure” research because there are no pure people. The questions we ask are influenced by our pasts and the world in which we grew up. The questions we ask are also influenced by the needs of the funder. And if the researcher goes ahead with findings that the funder doesn’t like, they are severely disciplined. That can be not winning the grants that keep you relevant at the university. Or if you actually work for the corporation, being fired.

And even when I was an unfunded researcher at university collaborating with police departments this mostly still applied. The part about the research being quashed was not an issue for me personally, but the *types* of questions asked are certainly influenced. A PD is unlikely to say ‘hey, lets examine some unintended consequences of my arrest policy’ – they are much more likely to say ‘hey, can you give me an argument to hire a few more guys?’. I do know of instances of others people work being limited from dissemination – the ones I am familiar with honestly it was stupid for the agencies to not let the researchers go ahead with the work, but I digress.

So we are all biased in some ways – we might as well admit it. What to do? One of my favorite passages in relation to our inherent bias is from Denis Wood’s introduction to his dissertation (see some more backstory via John Krygier). But here are some snippets from Wood’s introduction:

There is much rodomontade in the social sciences about being objective. Such talk is especially pretentious from the mouths of those whose minds have never been sullied by even the merest passing consideration of what it is that objectivity is supposed to be. There are those who believe it to consist in using the third person, in leaning heavily on the passive voice, in referring to people by numbers or letters, in reserving one’s opinion, in avoiding evaluative adjectives or adverbs,

ad nauseum. These of course are so many red herrings.

So we cannot be objective, no point denying it. But a few paragraphs later from Wood:

Yet this is no opportunity for erecting the scientific tombstone. Not quite yet. There is a pragmatic, possible, human out: Bare yourself.

Admit your attitudes, beliefs, politics, morals, opinions, enthusiasms, loves, odiums, ethics, religion, class, nationality, parentage, income, address, friends, lovers, philosophies, language, education. Unburden yourself of your secrets. Admit your sins. Let the reader decide if he would buy a used car from you, much less believe your science. Of course, since you will never become completely self-aware, no more in the subjective case than in the objective, you cannot tell your reader all. He doesn’t need it all. He needs enough. He will know.

This dissertation makes no pretense at being objective, whatever that ever was. I tell you as much as I can. I tell you as many of my beliefs as you could want to know. This is my Introduction. I tell you about this project in value-loaded terms. You will not need to ferret these out. They will hit you over the head and sock you in the stomach. Such terms, such opinions run throughout the dissertation. Then I tell you the story of this project, sort of as if you were in my – and not somebody else’s – mind. This is Part II of the dissertation. You may believe me if you wish. You may doubt every word. But I’m not conning you. Aside from the value-loaded vocabulary – when I think I’ve done something wonderful, or stupid, I don’t mind giving myself a pat on the back, or a kick in the pants. Parts I and II are what sloppy users of the English language might call “objective.” I don’t know about that. They’re conscientious, honest, rigorous, fair, ethical, responsible – to the extent, of course, that I am these things, no farther.

I think I’m pretty terrific. I tell you so. But you’ll make up your mind about me anyway. But I’m not hiding from you in the the third person passive voice – as though my science materialized out of thin air and marvelous intentions. I did these things. You know me, I’m

Denis Wood

We will never be able to scrub ourselves clean to be entirely objective – a pure researcher as Loomis puts its. But we can be transparent about the work we do, and let readers decide for themselves whether the work we bring forth is sufficient to overcome those biases or not.

]]>My question is from our CPP project on business improvement districts (Piza, Wheeler, Connealy, Feng 2020). The article indicates that you ran three of the microsynth matching variables as an average over each instead of the cumulative sum (street length, percent new housing structures, percent occupied structures). How did you get R to read the variables as averages instead of the entire sum of the treatment period of interest? I have the microsynth code you used to generate our models, but cannot seem to determine how you got R to read the variables as averages.

So Nate is talking about this paper, *Crime control effects of a police substation within a business improvement district: A quasi-experimental synthetic control evaluation* (Piza et al., 2020), and here is the balance table in the paper:

To be clear to folks, I did not balance on the averages, but simply reported the table in terms of averages. So here is the original readout from R:

So I just divided those noted rows by 314 to make them easier to read. You could divide values by the total number of treated units though in the original data to have microsynth match on the averages instead if you wanted to. Example below (this is R code, see the microsynth library and paper by Robbins et al., 2017):

```
library(microsynth)
#library(ggplot2) #not loading here, some issue
set.seed(10)
data(seattledmi) #just using data in the package
cs <- seattledmi
# calculating proportions
cs$BlackPerc <- (cs$BLACK/cs$TotalPop)*100
cs$FHHPerc <- (cs$FEMALE_HOU/cs$HOUSEHOLDS)*100
# replacing 0 pop with 0
cs[is.na(cs)] <- 0
cov.var <- c("TotalPop","HISPANIC","Males_1521","FHHPerc","BlackPerc")
match.out <- c("i_felony", "i_misdemea")
sea_prop <- microsynth(cs,
idvar="ID", timevar="time", intvar="Intervention",
start.pre=1, end.pre=12, end.post=16,
match.out.min=match.out,match.out=FALSE,
match.covar=FALSE,check.feas=FALSE,
match.covar.min=cov.var,
result.var=match.out)
summary(sea_prop) # balance table
```

And here you can see that we are matching on the cumulative sums for each of the areas, but we can divide our covariates by the number of treated units, and we will match on the proportional values.

```
# Can divide by 39 and get the same results
cs[,cov.var] <- cs[,cov.var]/39
sea_div <- microsynth(cs,
idvar="ID", timevar="time", intvar="Intervention",
start.pre=1, end.pre=12, end.post=16,
match.out.min=match.out,match.out=FALSE,
match.covar=FALSE,check.feas=FALSE,
match.covar.min=cov.var,
result.var=match.out)
summary(sea_div) # balance table
```

Note that these do not result in the same weights. If you look at the results you will see the treatment effects are slightly different. Also if you do:

```
# Showing weights are not equal
all.equal(sea_div$w$Weights,sea_prop$w$Weights)
```

It does not return True. Honestly not familiar enough with the procedure that microsynth uses to do the matching (Raking survey weights) to know if this is due to stochastic stuff or due to how the weighting algorithm works (I would have thought a linear change does not make a difference, but I was wrong).

On the bucket list is to do a matching algorithm that returns geographically contiguous areas and gives the weights all values of 1 (so creates comparable neighborhoods), instead of estimating Raking weights. That may be 5 years though before I get around to that. Gio has a nice map to show the way the weights work now is they may be all over the place (Circo et al., 2021) – I am not sure that is a good thing though.

But I did want to share some functions I used for the paper I worked with Nate on. First, this is for if you use the permutation approach, the function `prep_synth`

returns some of the data in a nicer format to make graphs and calculate your own stats:

```
# Function to scoop up the data nicely
prep_synth <- function(mod){
#Grab the plot data
plotStats <- mod[['Plot.Stats']]
#For the left graph
Treat <- as.data.frame(t(plotStats$Treatment))
Treat$Type <- "Treat"
#This works for my data at years, will not
#Be right for data with more granular time though
Treat$Year <- as.integer(rownames(Treat))
Cont <- as.data.frame(t(plotStats$Control))
Cont$Type <- "Control"
Cont$Year <- as.integer(rownames(Cont))
AllRes <- rbind(Treat,Cont)
#For the right graph
Perm <- as.data.frame(t(as.data.frame(plotStats$Difference)))
SplitStr <- t(as.data.frame(strsplit(rownames(Perm),"[.]")))
colnames(SplitStr) <- c("Type","Year")
rownames(SplitStr) <- 1:nrow(SplitStr)
SplitStr <- as.data.frame(SplitStr)
Perm$Type <- as.character(SplitStr$Type)
Perm$Year <- as.integer(as.character(SplitStr$Year))
Perm$Group <- ifelse(Perm$Type == 'Main','Treatment Effect','Permutations')
#Reordering factor levels for plots
AllRes$Type <- factor(AllRes$Type,levels=c('Treat','Control'))
levels(AllRes$Type) <- c('Treated','Synthetic Control')
Perm$Group <- factor(Perm$Group,levels=c('Treatment Effect','Permutations'))
#Exporting result
Res <- vector("list",length=2)
Res[[1]] <- AllRes
Res[[2]] <- Perm
names(Res) <- c("AggOutcomes","DiffPerms")
return(Res)
}
```

It works for the prior tables, but I really made these functions to work with when you used permutations to get the errors. (In the micro synth example, it is easier to work with permutations than in the state level example for synth, in which I think conformal prediction intervals makes more sense, see De Biasi & Circo, 2021 for a recent real example with micro place based data though.)

```
# Takes like 1.5 minutes
sea_perm <- microsynth(seattledmi,
idvar="ID", timevar="time", intvar="Intervention",
start.pre=1, end.pre=12, end.post=16,
match.out.min=match.out,match.out=FALSE,
match.covar=FALSE,check.feas=FALSE,
match.covar.min=cov.var,
result.var=match.out, perm=99)
res_prop <- prep_synth(sea_perm)
print(res_prop)
```

So the dataframe in the first slot is the overall treatment effect, and the second dataframe is a nice stacked version for the permutations. First, I really do not like the percentage change (see Wheeler, 2016 for the most direct critique, but I have a bunch on this site). So I wrote code to translate the treatment effects into crime count reductions instead of the percent change stuff.

```
# Getting the observed treatment effect on count scale
# vs the permutations
agg_fun <- function(x){
sdx <- sd(x)
minval <- min(x)
l_025 <- quantile(x, probs=0.025)
u_975 <- quantile(x, probs=0.975)
maxval <- max(x)
totn <- length(x)
res <- c(sdx,minval,l_025,u_975,maxval,totn)
return(res)
}
treat_count <- function(rp){
# Calculating the treatment effect based on permutations
keep_vars <- !( names(rp[[2]]) %in% c("Year","Group") )
out_names <- names(rp[[2]])[keep_vars][1:(sum(keep_vars)-1)]
loc_dat <- rp[[2]][,keep_vars]
agg_treat <- aggregate(. ~ Type, data = loc_dat, FUN=sum)
n_cols <- 2:dim(agg_treat)[2]
n_rows <- 2:nrow(agg_treat)
dif <- agg_treat[rep(1,max(n_rows)-1),n_cols] - agg_treat[n_rows,n_cols]
dif$Const <- 1
stats <- aggregate(. ~ Const, data = dif, FUN=agg_fun)
v_names <- c("se","min","low025","up975","max","totperm")
long_stats <- reshape(stats,direction='long',idvar = "Const",
varying=list(2:ncol(stats)),
v.names=v_names, times=out_names)
# Add back in the original stats
long_stats <- long_stats[,v_names]
rownames(long_stats) <- 1:nrow(long_stats)
long_stats$observed <- t(agg_treat[1,n_cols])[,1]
long_stats$outcome <- out_names
ord_vars <- c('outcome','observed',v_names)
return(long_stats[,ord_vars])
}
treat_count(res_prop)
```

So that is the cumulative total effect of the intervention. This is more similar to the WDD test (Wheeler & Ratcliffe, 2018), but since the pre-time period is matched perfectly, just is the differences in the post time periods. And here it uses the permutations to estimate the error, not any Poisson approximation.

But I often see folks concerned about the effects further out in time for synthetic control studies. So here is a graph that just looks at the instant effects for each time period, showing the difference via the permutation lines:

```
# GGPLOT graphs, individual lines
library(ggplot2)
perm_data <- res_prop[[2]]
# Ordering factors to get the treated line on top
perm_data$Group <- factor(perm_data$Group, c("Permutations","Treatment Effect"))
perm_data$Type <- factor(perm_data$Type, rev(unique(perm_data$Type)))
pro_perm <- ggplot(data=perm_data,aes(x=Year,y=i_felony,group=Type,color=Group,size=Group)) +
geom_line() +
scale_color_manual(values=c('grey','red')) + scale_size_manual(values=c(0.5,2)) +
geom_vline(xintercept=12) + theme_bw() +
labs(x=NULL,y='Felony Difference from Control') +
scale_x_continuous(minor_breaks=NULL, breaks=1:16) +
scale_y_continuous(breaks=seq(-10,10,2), minor_breaks=NULL) +
theme(panel.grid.major = element_line(linetype="dashed"), legend.title= element_blank(),
legend.position = c(0.2,0.8), legend.background = element_rect(linetype="solid", color="black")) +
theme(text = element_text(size=16), axis.title.y=element_text(margin=margin(0,10,0,0)))
```

And I also like looking at this for the cumulative effects as well, which you can see with the permutation lines widen over time.

```
# Cumulative vs Pointwise
perm_data$csum_felony <- ave(perm_data$i_felony, perm_data$Type, FUN=cumsum)
pro_cum <- ggplot(data=perm_data,aes(x=Year,y=csum_felony,group=Type,color=Group,size=Group)) +
geom_line() +
scale_color_manual(values=c('grey','red')) + scale_size_manual(values=c(0.5,2)) +
geom_vline(xintercept=12) + theme_bw() +
labs(x=NULL,y='Felony Difference from Control Cumulative') +
scale_x_continuous(minor_breaks=NULL, breaks=1:16) +
scale_y_continuous(breaks=seq(-20,20,5), minor_breaks=NULL) +
theme(panel.grid.major = element_line(linetype="dashed"), legend.title= element_blank(),
legend.position = c(0.2,0.8), legend.background = element_rect(linetype="solid", color="black")) +
theme(text = element_text(size=16), axis.title.y=element_text(margin=margin(0,10,0,0)))
```

If you do a ton of permutations (say 999 instead of 99), it would likely make more sense to do a fan chart type error bars and show areas of different percentiles instead of each individual line (Yim et al., 2020).

I will need to slate a totally different blog post to discuss instant vs cumulative effects for time series analysis. Been peer-reviewing quite a few time series analyses of Covid and crime changes – most everyone only focuses on instant changes, and does not calculate cumulative changes. See for example estimating excess deaths for the Texas winter storm power outage (Aldhous et al., 2021). Folks could do similar analyses for short term crime interventions. Jerry has a good example of using the Causal Impact package to estimate cumulative effects for a gang takedown intervention (Ratcliffe et al., 2017) for one criminal justice example I am familiar with.

Again for folks feel free to ask me anything. I may not always be able to do as deep a dive as this, but always feel free to reach out.

- Aldhous, P., Lee S.M., & Hirji, Z. (2021). The Texas Winter Storm And Power Outages Killed Hundreds More People Than The State Says.
*Buzzfeed*5/26/2021. - Circo, G. M., Krupa, J. M., McGarrell, E., & De Biasi, A. (2021). Focused Deterrence and Program Fidelity: Evaluating the Impact of Detroit Ceasefire.
*Justice Evaluation Journal*, 4(1), 112-130. - De Biasi, A., & Circo, G. (2021). Capturing Crime at the Micro-place: A Spatial Approach to Inform Buffer Size.
*Journal of Quantitative Criminology*, 37(2), 393-418. - Piza, E. L., Wheeler, A. P., Connealy, N. T., & Feng, S. Q. (2020). Crime control effects of a police substation within a business improvement district: A quasi-experimental synthetic control evaluation.
*Criminology & Public Policy*, 19(2), 653-684. - Ratcliffe, J. H., Perenzin, A., & Sorg, E. T. (2017). Operation Thumbs Down: A quasi-experimental evaluation of an FBI gang takedown in South Central Los Angeles.
*Policing: An International Journal of Police Strategies & Management*40(2), 442-458. - Robbins MW, Saunders J, Kilmer B (2017). A framework for synthetic control methods with high-dimensional, micro-level data: Evaluating a neighborhood-specific crime intervention,
*Journal of the American Statistical Association*, 112(517), 109-126. - Wheeler, A. P. (2016). Tables and graphs for monitoring temporal crime trends: Translating theory into practical crime analysis advice.
*International Journal of Police Science & Management*, 18(3), 159-172. - Wheeler, A. P., & Ratcliffe, J. H. (2018). A simple weighted displacement difference test to evaluate place based crime interventions.
*Crime Science*, 7(1), 1-9. - Yim, H. N., Riddell, J. R., & Wheeler, A. P. (2020). Is the recent increase in national homicide abnormal? Testing the application of fan charts in monitoring national homicide trends over time.
*Journal of Criminal Justice*, 66, 101656.