Exploratory Data Analysis (EDA) and Regression
This tutorial demonstrates some of the capabilities of R for exploring
two (or more) quantitative variables.
Bivariate exploratory data analysis
We begin by loading the Hipparcos dataset used in the
descriptive statistics tutorial, found at
hip <- read.table("http://astrostatistics.psu.edu/datasets/HIP_star.dat",
In the descriptive statistics tutorial,
we considered boxplots, a one-dimensional plotting technique.
We may perform a slightly more sophisticated analysis using boxplots
to get a glimpse at some bivariate structure. Let us
examine the values of Vmag, with objects
broken into categories according to the B minus V variable:
notch=T, varwidth=T, las=1, tcl=.5,
xlab=expression("B minus V"),
main="Can you find the red giants?",
cex=1, cex.lab=1.4, cex.axis=.8, cex.main=1)
axis(2, labels=F, at=0:12, tcl=-.25)
The notches in the boxes, produced using "notch=T", can be used to test for differences in
the medians (see
for details). With "varwidth=T", the box widths are proportional to the square roots of
the sample sizes. The "cex" options all give scaling factors, relative to default:
"cex" is for plotting text and symbols, "cex.axis" is for axis annotation,
"cex.lab" is for the x and y labels, and
"cex.main" is for main titles.
commands are used to add an axis to the current plot. The first such command
above adds smaller tick marks at all integers, whereas the second one adds the
axis on the right.
in the plot above are telling us something about the bivariate relationship
between the two variables. Yet it is probably easier to grasp this relationship
by producing a
The above plot looks too busy because of the default plotting character, set let
us use a different one:
Let's now use exploratory scatterplots to locate the
Hyades stars. This open cluster
should be concentrated both in the sky coordinates RA and DE, and also in the proper
motion variables pm_RA and pm_DE. We start by noticing a concentration of stars in the
See the cluster of stars with RA between 50 and 100 and with DE between
0 and 25?
Let's construct a logical (TRUE/FALSE) variable that will select only those stars
in the appropriate rectangle:
filter1 <- (RA>50 & RA<100 & DE>0 & DE<25)
Next, we select the proper motions according to those stars in the defined
rectangle. (As our cuts through the data
are parallel to the axes, this variable-by-variable classification approach
is sometimes called Classification and Regression
Trees or CART, a very common multivariate classification procedure.)
Let's replot after zooming in on the rectangle shown in red.
filter2 <- (pmRA>90 & pmRA<130 & pmDE>-60 & pmDE< -10) # Space in 'pmDE< -10' is necessary!
filter <- filter1 & filter2
Let's have a final look at the stars we have
identified using the
command to produce all bivariate plots for pairs of variables.
We'll exclude the first and fifth columns (the HIP identifying number
and the parallax, which is known to lie in a narrow band by construction).
Notice that indexing a matrix or vector using negative integers has the effect of
excluding the corresponding entries.
We see that there is one outlying star in the e_Plx variable,
indicating that its measurements are not reliable. We exclude this point:
filter <- filter & (e_Plx<5)
How many stars have we identified? The filter variable,
a vector of TRUE and FALSE, may be summed to reveal the
number of TRUE's (summation causes R to coerce the logical values to 0's and 1's).
As a final look at these data, let's consider the HR plot of Vmag versus B.V
but make the 92 Hyades stars we just identified look bigger (pch=20 instead of 46) and
color them red (col=2 instead of 1).
This shows the Zero Age Main Sequence, plus four
red giants, with great precision.
Linear and polynomial regression
Here is how one may reproduce the output seen in the regression lecture,
i.e., a linear regression relating BminusV to logL, where logL is the luminosity,
defined to be (15 - Vmag - 5 log(Plx)) / 2.5.
However, we'll use a different subset
of the data than the one seen in the lecture, namely, the main-sequence
mainseqhyades <- filter & (Vmag>4 | B.V<0.2)
logL <- (15 - Vmag - 5 * log10(Plx)) / 2.5
x <- logL[mainseqhyades]
y <- B.V[mainseqhyades]
regline <- lm(y~x)
abline(regline, lwd=2, col=2)
Note that the regression line passes exactly through the point (xbar, ybar):
points(mean(x), mean(y), col=3, pch=20, cex=3)
Here is a regression of y on exp(-x/4):
newx <- exp(-x/4)
regline2 <- lm(y~newx)
xseq <- seq(min(x), max(x), len=250)
lines(xseq, regline2$coef %*% rbind(1, exp(-xseq/4)), lwd=2, col=3)
For an implementation of the ridge regression technique mentioned in
lecture, see the
function in the MASS package. To use this function, you must
first type library(MASS). For a package that implements LASSO (which uses an
L1-penalty instead of the L2-penalty of ridge regression), check out, e.g.,
the lasso2 package on CRAN.
Let's consider a new dataset, one that comes from NASA's Swift satellite.
The statistical problem at hand is
modeling the X-ray afterglow of gamma ray
bursts. First, read in the dataset:
grb <- read.table("http://astrostatistics.psu.edu/datasets/GRB_afterglow.dat",
The skip=1 option in the previous statement tells R to ignore the first
row in the data file. You will see why this is necessary if you
look at the
Let's focus on the first two columns, which are times and X-ray fluxes:
This plot is very hard to interpret because of the scales, so let's take the log of each
x <- log(grb[,1])
y <- log(grb[,2])
plot(x,y,xlab="log time",ylab="log flux")
The relationship looks roughly linear, so let's try a linear model
(lm in R):
model1 <- lm(y~x)
abline(model1, col=2, lwd=2)
The "response ~ predictor(s)" format seen above is used
for model formulas in functions like
The model1 object just created is an object of
class "lm". The class of
an object in R can help to determine how it is treated by functions such as
model1 # same as print(model1)
Notice the sigma-hat, the R-squared and adjusted R-squared,
and the standard errors of the beta-hats in the output from
the summary function.
There is a lot of information contained in model1 that is not displayed by
For instance, we will use the model1$fitted.values and model1$residuals
information later when we look at some residuals plots.
Notice that the coefficient estimates
are listed in a regression table, which is standard regression output for
any software package. This table gives not only the estimates but their
standard errors as well, which enables us to determine whether the estimates
are very different from zero. It is possible to give individual confidence
intervals for both the intercept parameter and the slope parameter based on
this information, but remember that a line really requires both a slope and
an intercept. Since our goal is really to estimate a line here, maybe it
would be better if we could somehow obtain a confidence "interval"
for the lines themselves.
This may in fact be accomplished. By viewing a line as a single two-dimensional
point in (intercept, slope) space, we set up a one-to-one correspondence between
all (nonvertical) lines and all points in two-dimensional space. It is
possible to obtain a two-dimensional confidence ellipse for the
(intercept, slope) points, which may then be mapped back into the set of
lines to see what it looks like.
Performing all the calculations necessary to do this is somewhat tedious,
but fortunately, someone else has already done it and made it available to all
R users through CRAN, the Comprehensive R Archive Network. The necessary
part of the "car" (companion to applied regression) package, which may
installed onto the V: drive (we don't have write access to the default
location where R packages are installed) as follows:
install.packages("car",lib="V:/") # lib=... is not always necessary!
You will have to choose a CRAN mirror as part of the installation process.
Once the car package is installed, its contents can be loaded into the current
R session using the
If all has gone well, there is now a new set of functions, along with
Here is a 95% confidence ellipse for the (intercept, slope) pairs:
Remember that each point on the boundary or in the interior of this ellipse
represents a line. If we were to plot all of these lines on the original scatterplot,
the region they described would be a 95% confidence band for the true regression
line. My advisor, Dave Hunter,
and I wrote a simple function to draw the borders of this band
on a scatterplot. You can see this function at
www.stat.psu.edu/~dhunter/R/confidence.band.r; to read it into R, use
In this dataset, the confidence band is so narrow that it's hard to see.
However, the borders of the band are not straight. You can see the curvature
much better when there are fewer points or more variation, as in:
tmpx <- 1:10
tmpy <- 1:10+rnorm(10) # Add random Gaussian noise
Also note that increasing the sample size increases the precision of the
estimated line, thus narrowing the confidence band. Compare the previous plot
with the one obtained by replicating tmpx and tmpy 25 times each:
tmpx25 <- rep(tmpx,25)
tmpy25 <- rep(tmpy,25)
A related phenomenon is illustrated if we are given a value of the
predictor and asked to predict the response. Two types of intervals
are commonly reported in this case: A prediction interval for an
individual observation with that predictor value, and a confidence
interval for the mean of all individuals with that predictor value. The former
is always wider than the latter because it accounts for not only the
uncertainty in estimating the true line
but also the individual variation around the true line. This phenomenon may
be illustrated as follows. Again, we use a toy data set here because the effect
is harder to observe on our astronomical dataset. As usual, 95% is the
default confidence level.
predict(lm(tmpy~tmpx), data.frame(tmpx=7), interval="prediction")
text(c(7,7,7), .Last.value, "P",col=4)
predict(lm(tmpy~tmpx), data.frame(tmpx=7), interval="conf")
text(c(7,7,7), .Last.value, "C",col=5)
Polynomial curve-fitting: Still linear regression!
Because there appears to be a bit of a bend in the scatterplot, let's try
fitting a quadratic curve instead of a linear curve. Note: Fitting a
quadratic curve is still considered linear regression. This may seem strange,
but the reason is that the quadratic regression model assumes that the response y
is a linear combination of 1, x, and x2. Notice the special
form of the
lm command when we implement quadratic
I function means "as is" and it
resolves any ambiguity in the model formula:
model2 <- lm(y~x+I(x^2))
Here is how to find the estimates of beta using the closed-form solution
seen in lecture:
X <- cbind(1, x, x^2) # Create nx3 X matrix
solve(t(X) %*% X) %*% t(X) %*% y # Compare to the coefficients above
Plotting the quadratic curve is not a simple matter of using the
abline function. To obtain
the plot, we'll first create a sequence of x values, then apply
the linear combination implied by the regression model using matrix
xx <- seq(min(x),max(x),len=200)
yy <- model2$coef %*% rbind(1,xx,xx^2)
Diagnostic residual plots
Comparing the (red) linear fit with the (green) quadratic fit visually, it
does appear that the latter looks slightly better. However, let's check some
diagnostic residual plots for these two models. To do this, we'll use the
plot.lm command, which is capable of
producing six different types of diagnostic plots. We will only consider
two of the six: A plot of residuals versus fitted values and a normal
quantile-quantile (Q-Q) plot.
It is not actually necessary to type
in the previous command;
plot would have
worked just as well. This is because model1 is an object of
class "lm" -- a
fact that can be verified by typing "class(model1)" -- and so R knows to apply
plot.lm if we
simply type "plot(model1, which=1:2)".
Looking at the first plot, residuals vs. fitted, we immediately see a problem
with model 1. A "nice" residual plot should have residuals both above and below
the zero line, with the vertical spread around the line roughly of the same
magnitude no matter what the value on the horizontal axis. Furthermore, there
should be no obvious curvature pattern. The red line is a
produced to help discern
any patterns (more on lowess later), but this line is not necessary in the
case of model1 to see the clear pattern of negative residuals on the left,
positive in the middle, and negative on the right. There is curvature here
that the model missed!
Pressing the return key to see the second plot reveals a normal quantile-quantile
plot. The idea behind this plot is that it will make a random sample from
a normal distribution look like a straight line. To the extent that the
normal Q-Q plot does not look like a straight line, the assumption of
normality of the residuals is suspicious. For model1, the clear S-shaped
pattern indicates non-normality of the residuals.
How do the same plots look for the quadratic fit?
These plots look much better. There is a little bit of waviness in the
residuals vs. fitted plot, but the pattern is nowhere near as obvious as it was
before. And there appear to be several outliers among the residuals on the
normal Q-Q plot, but the normality assumption looks much less suspect here.
The residuals we have been using in the above plots are the ordinary
residuals. However, it is important to keep in mind that even if all
of the assumptions of the regression model are perfectly true (including
the assumption that all errors have the same variance), the variances of
the residuals are not equal. For this reason, it is better to use
the studentized residuals. Unfortunately, R reports the ordinary residuals
by default and it is necessary to call another function to obtain the
The good news is that in most datasets,
residual plots using the studentized residuals are essentially indistinguishable
in shape from residual plots using the ordinary residuals, which means
that we would come to the same conclusions regardless of which set
of residuals we use.
rstu <- rstudent(model2)
To see how similar the studentized residuals are to a scaled version of
the ordinary residuals (called the standardized residuals), we can depict both
on the same plot:
rsta <- rstandard(model2)
points(model2$fit, rsta, col=2, pch=3)
Collinearity and variance inflation factors
Let's check the variance inflation factors (VIFs) for the quadratic fit.
The car package that we installed earlier contains a function called
vif that does this automatically. Check its help page by
typing "?vif" if you wish.
Note that it does not make sense to look at variance inflation factors
for model1, which has only one term (try it and see what happens). So we'll
start by examining model2.
of more than 70 indicate a high degree of collinearity between the values of
x and x^2 (the two predictors). This is not surprising, since x has a range
from about 5 to 13. In fact, it is easy to visualize
the collinearity in a plot:
plot(x,x^2) # Note highly linear-looking plot
To correct the collinearity, we'll replace x and x^2 by (x-m) and (x-m)^2,
where m is the sample mean of x:
centered.x <- x-mean(x)
model2.2 <- lm(y ~ centered.x + I(centered.x^2))
This new model has much lower VIFs, which means that we have greatly
reduced the collinearity. However, the fit
is exactly the same: It is still the best-fitting quadratic curve. We may
demonstrate this by plotting both fits on the same set of axes:
plot(x,y,xlab="log time",ylab="log flux")
yy2 <- model2.2$coef %*% rbind(1, xx-mean(x), (xx-mean(x))^2)
lines(xx, yy, lwd=2, col=2)
lines(xx, yy2, lwd=2, col=3, lty=2)
Model selection using AIC and BIC
Let's compare the AIC and BIC values for the linear and the quadratic fit.
Without getting too deeply into details,
the idea behind these criteria is that we know the model with more parameters
(the quadratic model) should achieve a higher maximized log-likelihood
than the model with fewer parameters
(the linear model). However, it may be that the additional increase in the
log-likelihood statistic achieved with more parameters is not worth adding
the additional parameters. We may test whether it is worth adding the additional
parameters by penalizing the log-likelihood by subtracting some positive
multiple of the number of parameters. In practice, for technical reasons
we take -2 times the log-likelihood, add a positive multiple of the number of
parameters, and look for the smallest resulting value. For AIC, the positive
multiple is 2; for BIC, it is the natural log of n, the number of observations.
We can obtain both the AIC and BIC results using the
AIC function. Remember that
R is case-sensitive, so "AIC" must be all capital letters.
The value of AIC for model2 is smaller than that for model1, which indicates
that model2 provides a better fit that is worth the additional parameters.
However, AIC is known to tend to overfit sometimes, meaning that it
sometimes favors models with more parameters than they should have. The
BIC uses a larger penalty than AIC, and it often seems to do a slightly
better job; however, in this case we see there is no difference in
n <- length(x)
It did not make any difference in the above output that we used model2 (with
the uncentered x values) instead of model2.2 (with the centered values). However,
if we had looked at the AIC or BIC values for a model containing ONLY the
quadratic term but no linear term, then we would see a dramatic difference.
Which one of the following would you expect to be higher (i.e., indicating
a worse fit), and why?
Other methods of curve-fitting
Let's try a nonparametric fit, given by
First we plot the linear (red) and quadratic (green) fits,
then we overlay the lowess fit in blue:
plot(x,y,xlab="log time",ylab="log flux")
abline(model1, lwd=2, col=2)
lines(xx, yy, lwd=3, col=3)
npmodel1 <- lowess(y~x)
lines(npmodel1, col=4, lwd=2)
It is hard to see the pattern of the lowess curve in the plot. Let's replot it with
no other distractions. Notice that the "type=n" option to
plot function causes the axes to
be plotted but not the points.
plot(x,y,xlab="log time",ylab="log flux", type="n")
lines(npmodel1, col=4, lwd=2)
This appears to be a piecewise linear curve. An analysis that assumes a piecewise
linear curve will be carried out on these data later in the week.
In the case of non-polynomial (but still parametric) curve-fitting,
we can use
If we replace the response y by the original (nonlogged) flux values, we might
posit a parametric model of the form flux = exp(a+b*x),
where x=log(time) as before.
Compare a nonlinear approach (in red) with a nonparametric
approach (in green) for this case:
flux <- grb[,2]
nlsmodel1 <- nls(flux ~ exp(a+b*x), start=list(a=0,b=0))
npmodel2 <- lowess(flux~x)
plot(x, flux, xlab="log time", ylab="flux")
lines(xx, exp(9.4602-.9674*xx), col=2, lwd=2)
lines(npmodel2, col=3, lwd=2)
Interestingly, the coefficients of the nonlinear least squares fit are different
than the coefficients of the original linear model fit on the logged data,
even though these coefficients have exactly the same interpretation: If
flux = exp(a + b*x), then shouldn't log(flux) = a + b*x?
The difference arises because these two fitting methods calculate (and subsequently
minimize) the residuals
on different scales. Try plotting exp(a + b*xx) on the scatterplot of x vs. flux
for both (a,b) solutions to see what happens. Next, try plotting a + b*xx on
the scatterplot of x vs. log(flux) to see what happens.
If outliers appear to have too large an influence over the least-squares
solution, we can also try resistant regression, using the
function in the MASS package.
The basic idea behind lqs is that the largest residuals (presumably
corresponding to "bad" outliers) are ignored.
The results for our log(flux) vs. log(time)
example look terrible but are
very revealing. Can you understand why the output from lqs looks so very
different from the least-squares output?
lqsmodel1 <- lqs(y~x, method="lts")
plot(x,y,xlab="log time",ylab="log flux")
Finally, let's consider least absolute deviation regression, which may be
considered a milder form of resistant regression than
lqs. In least absolute deviation
regression, even large residuals have an influence on the regression line
lqs), but this influence
is less than in least squares regression.
To implement it, we'll
use a function called rq
in the "quantreg" package. Like the "car" package, this package is
not part of the standard distribution of R, so we'll need to download it.
Also, we will need to install the "SparseM" package as this is required for the
"quantreg" package. In order to do this, we must tell R where to store the installed library
install.packages(c("SparseM","quantreg"),lib="V:/") # lib=... is not always necessary!
Assuming the quantreg package is loaded, we may now compare the least-squares fit (red)
with the least absolute deviations fit (green). In this example, the two
fits are nearly identical:
rqmodel1 <- rq(y~x)
plot(x,y,xlab="log time",ylab="log flux")