# Data analysis Ben Graham October 13, 2015 MA930, University of Warwick

```Data analysis
Ben Graham
MA930, University of Warwick
October 13, 2015
Ch4: Joint and Marginal Distributions
Def 4.1.1nAn n-dimensional random vector
is a function
n
X = (Xi )i from sample space S to R .
n = 2; roll two dice
=1
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rnorm(10)
rbinom(10,5,0.5)
Discrete case: Joint p.m.f. fX (x , . . . , xn):
1
P[X ∈ A] =
X
x =(x1 ,...,xn )∈A
fX (x )
Continuous case: Joint p.d.f. fX (x , . . . , xn):
1
P[X ∈ A] =
ˆ
x =(x1 ,...,xn )∈A
fX (x )dx
1
. . . dxn
Marginal distributions
Discrete case: pmf fX Y : R
,
fX (x ) =
2
→R
.
X
y :fX ,Y (x ,y )&gt;0
Continuous case: pdf fX Y : R
,
fX (x ) =
2
ˆ
fX ,Y (x , y )
.
→R
fX ,Y (x , y )dy
Example
Discrete
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X , Y ∈ {1, . . . , 6} independent dice rolls
Z =X +Y
p.m.f. fX ,Y
p.m.f. fX ,Z
Example
Continuous
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X , Y ∼ N (0p, 1) i.i.d.r.v
Z = ρX + 1 − ρ Y ∼ N (0, 1)
fX ,Y (x , y ) = fX (x )fY (y )
fX ,Z (x , z ) = fX (x )fZ |X (z | x ) [(Z | X ) ∼ N (ρX , 1 − ρ
2
bivariate normal distribution
2
)]
4.2 Conditional Distributions and Independence
Random variables X and Y are independent if
for all x , y , {X &lt; x } and {Y &lt; y } are independent
fX Y (x , y ) = fX (x )fY (y ) [continuous p.d.f.s or discrete p.m.f.s]
φX Y (t ) = φX (t )φY (t ) [characteristic functions]
Examples
X , Y ∼ Bernoulli (1/2)
X , Y ∼ N (0, 1)
Independence →Covariance=0
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,
+
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E[XY ] = E[X ]E[Y ]
Covariance=0 6→independence
X ∼ N (0, 1)
Y ∈ {−1, +1} independent of X
Cov(X,XY)=0
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Sums of Normal distributions
Example 4.3.4.
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X ∼ N (&micro;X , σX )
Y ∼ N (&micro;Y , σY )
X , Y independent
Then X + Y ∼ N (&micro;X + &micro;Y , σX + σY ).
2
2
2
Random sample
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X
2
, . . . , Xn ∼ N (&micro;, σ 2 )
P
2
i Xi ∼ N (n&micro;, nσ )
Then
1
I
X̄
=
Z
=
X Xi
i
I
n
∼ N (&micro;, σ 2 /n)
X Xi − &micro;
√ ∼ N( , )
σ n
i
01
Sums of Poissons
Theorem 4.3.2
X ∼Poisson(θ)
Y ∼Poisson(λ)
X , Y independent.
Then X + Y ∼Poisson(θ + λ)
Ex 4.4.1 Conversely
Y ∼Poisson(λ)
X | Y ∼Bin(Y , p)
Then X ∼Poisson(λp)
Random Samples
X , . . . , Xn ∼Poisson(λ) i.i.d.r.v.
P
i Xi ∼Poisson(nλ) ≈ N (nλ, nλ)
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1
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X̄
∼≈ N (λ, λ/n)
Covariance and correlation
Def 4.5.1 Covariance:
Cov (X , Y ) = E [(X − EX )(Y − EY )] = E[XY ] − EX EY
Def 4.5.2 Correlation
Cov (X , Y )
Var (X )Var (Y )
Thm 4.5.6 If X and Y are r.v. and a, b ∈ R, then
ρXY = p
Var (aX + bY ) = a Var (X ) + b Var (Y ) + 2abCov (X , Y )
2
Special case X , Y independent.
2
Def 4.5.10 Bivariate distribution
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&micro;X , &micro;Y ∈ R
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σx , σY &gt;
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pdf
0
fX ,Y (x , y )
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Or pdf
1
2
1
exp 2 1
2
p
− ρ2 )−1
= ( πσX σY
x − &micro; 2
x
&times;
( − ρ2 )
σX
x − &micro; Y − &micro; Y − &micro; 2 x
Y +
Y
− ρ
σX
σY
σY
f (x ) = p
k = 2, x
1
exp
(2π)k |Σ|
∈ Rk , &micro; =
&micro;x
&micro;Y
− (x − &micro;)Σ−1 (x − &micro;)
, Σ=
σX2
ρσX σY
ρσX σY
σY2
Def 4.6.2 Multinomial distribution
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m trials / repeat events
n possible outcomes, probabilities p , . . . , pn sum to one.
Let xi ∈ {0, 1, . . . , n} count the number of type-i outcomes
1
joint pmf
f (x
1
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, . . . , xn ) =
x
1
m!
! . . . xn !
px1 . . . pnxn
1
where X xi = m
i
Negative correlations Cov (Xi , Xj ) = −mpi pj (i 6= j )
Modeling tables of categorical variables.
Ch5 Random samples
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X , . . . , Xn is called a random sample of size n from population
f (x ) if they are independent, identically distributed random
variables (i.i.d.r.v.) with marginal distribution function f (x ).
1
Think of them as being a random sample from a population
that is much larger than n. The probability distribution
represents the true distribution of values in the larger
population.
Condorcet's Jury Principle:
plot(sapply(seq(0, 1, 0.01), function(p) pbinom(6, 12, p)))
Getting representative samples can be hard, i.e. telephone
surveys.
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Does everyone have a landline?
Does everyone choose to talk to strangers phoning you up
during dinner?
Are people shy about expressing some preferences?
http://www.bbc.co.uk/news/uk-politics-33228669
Parameters
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Parameter θ controlling the distribution f (x | θ)
Frequentist statistics:
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Bayesian:
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θ is xed but unknown
Choose θ̂ = θ̂(data) to estimate θ.
Joint distribution f (θ)f (x | θ)
f (θ) is the prior distribution
Example.
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Exponential f (x | θ) = θ exp(−θ
Q x ).
P
Joint distribution f (x | θ) = =1 f (x | θ) = θ exp(−θ
N.B. f (x | θ)only depends on the x via their sum.
n
n
i
i
i
i
x)
i
Statistics
Def 5.2.1
Let X , . . . , Xn ∼ f (x | θ)
Let T (x , . . . , xn) denote some function of the data,
i.e. T : Rn → R.
T is a statistic
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1
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T (x1 , . . . , x
T (x1 , . . . , x
n
n
) = x1
) = max
mean, median, etc.
i
x
i
, Var(X ), etc, are not statistics.
The probability distribution of T is called the sampling
distribution of T .
θ,EX
Common statistics
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i.i.d.r.v.
Sample mean
X
1
, . . . , Xn
X̄
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=
1
+ &middot; &middot; &middot; + Xn
n
=
n
X
i =1
Xi
Sample variance
&quot; n
#
n
X
X
1
1
S =
(Xi − X̄ ) =
Xi − nX̄
n−1 i
n−1 i
If the mean EXi and variance Var(Xi ) exist, then these are
unbiased estimates.
2
2
=1
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X
2
=1
2
*Def 5.4.1 Order Statistics
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The order statistics of a random sample X , . . . , Xn are the
values placed into increasing order: X , . . . , X n .
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(1)
X(1) = min X
X(2) =second smallest
i
( )
i
...
X( ) = max X
The sample range is R = X n − X
The sample median is
n
i
i
( )
M=
(1)
X((n+ )/ )
n odd
X(n/ ) + X(n/ + ) n even
(
1
2
2
1
1
2
2
2
1
The median may give a better sense of what is typical than the
mean.
Quantiles
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For p ∈ [0, 1], the p quantile is (R, method 7 of 9)
(1−γ)x j +γ x j , (n−1)p &lt; j ≤ (n−1)p +1, γ = np +1−p −j
Trimmed/truncated mean
()
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p ∈ [0, 1]
( +1)
Remove the smallest and biggest np items from the sample
Take the mean of what is left
i.e. LIBOR, 18 banks, top and bottom 4 removed
Cauchy location parameter
Theorem 5.4.4 Distribution of the order statistics
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Sample of size n with c.d.f. FX and pdf fX
Binomial distribution:
n X
n
FX j (x ) =
[F (x )]k [1 − FX (x )]n
k X
( )
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Dierentiate:
fX(j ) (x ) =
−k
k =j
n!
fX (x ) [FX (x )]j −
(j − 1)!(n − j )!
1
1
[ − FX (x )]n−j .
More sample mean
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Characteristic function
φX̄ (t ) = [φX (t /n)]n ≈
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+ o (t /n)
Thm 5.3.1: If Xi ∼ N (&micro;, σ ), then
2
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1 + it EnX
X̄ and S 2 are independent
X̄ ∼ N (&micro;, σ2 /n).
(n − 1)S 2 /σ 2 ∼ χ2−1
Proof: For simplicity, assume &micro; = 0 and σ = 1.
n
2
n
Ingredients for the proof
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If e , . . . , en form
an orthonormal basis for Rn.
n
Then (ei &middot; X )i are i.i.d.r.v N (0, 1).
Let e = ( n , . . . , n )
The Γ(α, β) distribution is dened
f (x ) = C (α, β)x exp(−β x ).
The χk distribution is the Γ(k /2, 1/2) distribution.
If X ∼ N (0, 1), then X ∼ χ
The sum of k independent χ r.v. is χk .
1
=1
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√1
1
√1
α−1
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2
2
2
1
2
1
2
Derived distributions
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If Z ∼ N (0, 1) and V ∼ χk then Z /pV /k ∼ tk (Student's t)
2
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X̄ −√&micro;
∼t
S / n n−
If U ∼ χk and V ∼ χ are independent, then UV k ∼ Fk
Suppose X , . . . , Xm ∼ N (&micro;X , σX ) and
Y , . . . , Yn ∼ N (&micro;Y , σY ). Then
2
/
/`
2
`
,`
2
1
1
2
SX /σX
SY /σY
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1
2
2
2
2
∼ Fm−1,n−1
These distributions are also used for linear regression/ANOVA.
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