Probability and Statistics Cookbook

Probability and Statistics
Cookbook
Copyright c Matthias Vallentin, 2011
vallentin@icir.org
12th December, 2011

This cookbook integrates a variety of topics in probability the-
ory and statistics. It is based on literature [1, 6, 3] and in-class
material from courses of the statistics department at the Uni-
versity of California in Berkeley but also influenced by other
sources [4, 5]. If you find errors or have suggestions for further
topics, I would appreciate if you send me an email. The most re-
cent version of this document is available at http://matthias.
vallentin.net/probability-and-statistics-cookbook/. To
reproduce, please contact me.
Contents
1 Distribution Overview 3
1.1 Discrete Distributions . . . . . . . . . . 3
1.2 Continuous Distributions . . . . . . . . 4
2 Probability Theory 6
3 Random Variables 6
3.1 Transformations . . . . . . . . . . . . . 7
4 Expectation 7
5 Variance 7
6 Inequalities 8
7 Distribution Relationships 8
8 Probability and Moment Generating
Functions 9
9 Multivariate Distributions 9
9.1 Standard Bivariate Normal . . . . . . . 9
9.2 Bivariate Normal . . . . . . . . . . . . . 9
9.3 Multivariate Normal . . . . . . . . . . . 9
10 Convergence 9
10.1 Law of Large Numbers (LLN) . . . . . . 10
10.2 Central Limit Theorem (CLT) . . . . . 10
11 Statistical Inference 10
11.1 Point Estimation . . . . . . . . . . . . . 10
11.2 Normal-Based Confidence Interval . . . 11
11.3 Empirical distribution . . . . . . . . . . 11
11.4 Statistical Functionals . . . . . . . . . . 11
12 Parametric Inference 11
12.1 Method of Moments . . . . . . . . . . . 11
12.2 Maximum Likelihood . . . . . . . . . . . 12
12.2.1 Delta Method . . . . . . . . . . . 12
12.3 Multiparameter Models . . . . . . . . . 12
12.3.1 Multiparameter delta method . . 13
12.4 Parametric Bootstrap . . . . . . . . . . 13
13 Hypothesis Testing 13
14 Bayesian Inference 14
14.1 Credible Intervals . . . . . . . . . . . . . 14
14.2 Function of parameters . . . . . . . . . . 14
14.3 Priors . . . . . . . . . . . . . . . . . . . 15
14.3.1 Conjugate Priors . . . . . . . . . 15
14.4 Bayesian Testing . . . . . . . . . . . . . 15
15 Exponential Family 16
16 Sampling Methods 16
16.1 The Bootstrap . . . . . . . . . . . . . . 16
16.1.1 Bootstrap Confidence Intervals . 16
16.2 Rejection Sampling . . . . . . . . . . . . 17
16.3 Importance Sampling . . . . . . . . . . . 17
17 Decision Theory 17
17.1 Risk . . . . . . . . . . . . . . . . . . . . 17
17.2 Admissibility . . . . . . . . . . . . . . . 17
17.3 Bayes Rule . . . . . . . . . . . . . . . . 18
17.4 Minimax Rules . . . . . . . . . . . . . . 18
18 Linear Regression 18
18.1 Simple Linear Regression . . . . . . . . 18
18.2 Prediction . . . . . . . . . . . . . . . . . 19
18.3 Multiple Regression . . . . . . . . . . . 19
18.4 Model Selection . . . . . . . . . . . . . . 19
19 Non-parametric Function Estimation 20
19.1 Density Estimation . . . . . . . . . . . . 20
19.1.1 Histograms . . . . . . . . . . . . 20
19.1.2 Kernel Density Estimator (KDE) 21
19.2 Non-parametric Regression . . . . . . . 21
19.3 Smoothing Using Orthogonal Functions 21
20 Stochastic Processes 22
20.1 Markov Chains . . . . . . . . . . . . . . 22
20.2 Poisson Processes . . . . . . . . . . . . . 22
21 Time Series 23
21.1 Stationary Time Series . . . . . . . . . . 23
21.2 Estimation of Correlation . . . . . . . . 24
21.3 Non-Stationary Time Series . . . . . . . 24
21.3.1 Detrending . . . . . . . . . . . . 24
21.4 ARIMA models . . . . . . . . . . . . . . 24
21.4.1 Causality and Invertibility . . . . 25
21.5 Spectral Analysis . . . . . . . . . . . . . 25
22 Math 26
22.1 Gamma Function . . . . . . . . . . . . . 26
22.2 Beta Function . . . . . . . . . . . . . . . 26
22.3 Series . . . . . . . . . . . . . . . . . . . 27
22.4 Combinatorics . . . . . . . . . . . . . . 27

1 Distribution Overview
1.1 Discrete Distributions
Notation1
FX (x) fX (x) E [X] V [X] MX (s)
Uniform Unif {a, . . . , b}



0 x < a
x −a+1
b−a
a ≤ x ≤ b
1 x > b
I(a < x < b)
b − a + 1
a + b
2
(b − a + 1)2
− 1
12
eas
− e−(b+1)s
s(b − a)
Bernoulli Bern (p) (1 − p)1−x
px
(1 − p)1−x
p p(1 − p) 1 − p + pes
Binomial Bin (n, p) I1−p(n − x, x + 1)
n
x
px
(1 − p)n−x
np np(1 − p) (1 − p + pes
)n
Multinomial Mult (n, p)
n!
x1! . . . xk!
px1
1 · · · p
xk
k
k
i=1
xi = n npi npi(1 − pi)
k
i=0
piesi
n
Hypergeometric Hyp (N, m, n) ≈ Φ
x − np
np(1 − p)
m
x
m−x
n−x
N
x
nm
N
nm(N − n)(N − m)
N2(N − 1)
Negative Binomial NBin (r, p) Ip(r, x + 1)
x + r − 1
r − 1
pr
(1 − p)x
r
1 − p
p
r
1 − p
p2
p
1 − (1 − p)es
r
Geometric Geo (p) 1 − (1 − p)x
x ∈ N+
p(1 − p)x−1
x ∈ N+ 1
p
1 − p
p2
p
1 − (1 − p)es
Poisson Po (λ) e−λ
x
i=0
λi
i!
λx
e−λ
x!
λ λ eλ(es
−1)
q q q q q q
Uniform (discrete)
x
PMF
a b
1
n
q q q q
q
q
q
q
q
q
q
q
q
q
q
q
q
q
q
q q q q q q q q q q q q q q q q q q q q q
0 10 20 30 40
0.000.050.100.150.200.25
Binomial
x
PMF
q n = 40, p = 0.3
n = 30, p = 0.6
n = 25, p = 0.9
q
q
q
q
q
q
q
q
q q
0 2 4 6 8 10
0.00.20.40.60.8
Geometric
x
PMF
q p = 0.2
p = 0.5
p = 0.8
q q
q
q
q
q q q q q q q q q q q q q q q q
0 5 10 15 20
0.00.10.20.3
Poisson
x
PMF
q λ = 1
λ = 4
λ = 10
1We use the notation γ(s, x) and Γ(x) to refer to the Gamma functions (see §22.1), and use B(x, y) and Ix to refer to the Beta functions (see §22.2).
3

1.2 Continuous Distributions
Notation FX (x) fX (x) E [X] V [X] MX (s)
Uniform Unif (a, b)



0 x < a
x−a
b−a
a < x < b
1 x > b
I(a < x < b)
b − a
a + b
2
(b − a)2
12
esb
− esa
s(b − a)
Normal N µ, σ2
Φ(x) =
x
−∞
φ(t) dt φ(x) =
1
σ
√
2π
exp −
(x − µ)2
2σ2
µ σ2
exp µs +
σ2
s2
2
Log-Normal ln N µ, σ2 1
2
+
1
2
erf
ln x − µ
√
2σ2
1
x
√
2πσ2
exp −
(ln x − µ)2
2σ2
eµ+σ2
/2
(eσ2
− 1)e2µ+σ2
Multivariate Normal MVN (µ, Σ) (2π)−k/2
|Σ|−1/2
e− 1
2
(x−µ)T
Σ−1
(x−µ)
µ Σ exp µT
s +
1
2
sT
Σs
Student’s t Student(ν) Ix
ν
2
,
ν
2
Γ ν+1
2
√
νπΓ ν
2
1 +
x2
ν
−(ν+1)/2
0 0
Chi-square χ2
k
1
Γ(k/2)
γ
k
2
,
x
2
1
2k/2Γ(k/2)
xk/2−1
e−x/2
k 2k (1 − 2s)−k/2
s < 1/2
F F(d1, d2) I d1x
d1x+d2
d1
2
,
d1
2
(d1x)d1 d
d2
2
(d1x+d2)d1+d2
xB d1
2
, d1
2
d2
d2 − 2
2d2
2(d1 + d2 − 2)
d1(d2 − 2)2(d2 − 4)
Exponential Exp (β) 1 − e−x/β 1
β
e−x/β
β β2 1
1 − βs
(s < 1/β)
Gamma Gamma (α, β)
γ(α, x/β)
Γ(α)
1
Γ (α) βα
xα−1
e−x/β
αβ αβ2 1
1 − βs
α
(s < 1/β)
Inverse Gamma InvGamma (α, β)
Γ α, β
x
Γ (α)
βα
Γ (α)
x−α−1
e−β/x β
α − 1
α > 1
β2
(α − 1)2(α − 2)2
α > 2
2(−βs)α/2
Γ(α)
Kα −4βs
Dirichlet Dir (α)
Γ k
i=1 αi
k
i=1 Γ (αi)
k
i=1
xαi−1
i
αi
k
i=1 αi
E [Xi] (1 − E [Xi])
k
i=1 αi + 1
Beta Beta (α, β) Ix(α, β)
Γ (α + β)
Γ (α) Γ (β)
xα−1
(1 − x)β−1 α
α + β
αβ
(α + β)2(α + β + 1)
1 +
∞
k=1
k−1
r=0
α + r
α + β + r
sk
k!
Weibull Weibull(λ, k) 1 − e−(x/λ)k k
λ
x
λ
k−1
e−(x/λ)k
λΓ 1 +
1
k
λ2
Γ 1 +
2
k
− µ2
∞
n=0
sn
λn
n!
Γ 1 +
n
k
Pareto Pareto(xm, α) 1 −
xm
x
α
x ≥ xm α
xα
m
xα+1
x ≥ xm
αxm
α − 1
α > 1
xα
m
(α − 1)2(α − 2)
α > 2 α(−xms)α
Γ(−α, −xms) s < 0
4

q q
Uniform (continuous)
x
PDF
a b
1
b − a
q q
−4 −2 0 2 4
0.00.20.40.60.8
Normal
x
φ(x)
µ = 0, σ2
= 0.2
µ = 0, σ2
= 1
µ = 0, σ2
= 5
µ = −2, σ2
= 0.5
0.0 0.5 1.0 1.5 2.0 2.5 3.0
0.00.20.40.60.81.0
Log−normal
x
PDF
µ = 0, σ2
= 3
µ = 2, σ2
= 2
µ = 0, σ2
= 1
µ = 0.5, σ2
= 1
µ = 0.25, σ2
= 1
µ = 0.125, σ2
= 1
−4 −2 0 2 4
0.00.10.20.30.4
Student's t
x
PDF
ν = 1
ν = 2
ν = 5
ν = ∞
0 2 4 6 8
0.00.10.20.30.40.5
χ2
x
PDF
k = 1
k = 2
k = 3
k = 4
k = 5
0 1 2 3 4 5
0.00.51.01.52.02.53.0
F
x
PDF
d1 = 1, d2 = 1
d1 = 2, d2 = 1
d1 = 5, d2 = 2
d1 = 100, d2 = 1
d1 = 100, d2 = 100
0 1 2 3 4 5
0.00.51.01.52.0
Exponential
x
PDF
β = 2
β = 1
β = 0.4
0 5 10 15 20
0.00.10.20.30.40.5
Gamma
x
PDF
α = 1, β = 2
α = 2, β = 2
α = 3, β = 2
α = 5, β = 1
α = 9, β = 0.5
0 1 2 3 4 5
01234
Inverse Gamma
x
PDF
α = 1, β = 1
α = 2, β = 1
α = 3, β = 1
α = 3, β = 0.5
0.0 0.2 0.4 0.6 0.8 1.0
0.00.51.01.52.02.53.0
Beta
x
PDF
α = 0.5, β = 0.5
α = 5, β = 1
α = 1, β = 3
α = 2, β = 2
α = 2, β = 5
0.0 0.5 1.0 1.5 2.0 2.5
0.00.51.01.52.02.5
Weibull
x
PDF
λ = 1, k = 0.5
λ = 1, k = 1
λ = 1, k = 1.5
λ = 1, k = 5
0 1 2 3 4 5
0123
Pareto
x
PDF
xm = 1, α = 1
xm = 1, α = 2
xm = 1, α = 4
5

2 Probability Theory
Deﬁnitions
• Sample space Ω
• Outcome (point or element) ω ∈ Ω
• Event A ⊆ Ω
• σ-algebra A
1. ∅ ∈ A
2. A1, A2, . . . , ∈ A =⇒
∞
i=1 Ai ∈ A
3. A ∈ A =⇒ ¬A ∈ A
• Probability Distribution P
1. P [A] ≥ 0 ∀A
2. P [Ω] = 1
3. P
∞
i=1
Ai =
∞
i=1
P [Ai]
• Probability space (Ω, A, P)
Properties
• P [∅] = 0
• B = Ω ∩ B = (A ∪ ¬A) ∩ B = (A ∩ B) ∪ (¬A ∩ B)
• P [¬A] = 1 − P [A]
• P [B] = P [A ∩ B] + P [¬A ∩ B]
• P [Ω] = 1 P [∅] = 0
• ¬( n An) = n ¬An ¬( n An) = n ¬An DeMorgan
• P [ n An] = 1 − P [ n ¬An]
• P [A ∪ B] = P [A] + P [B] − P [A ∩ B]
=⇒ P [A ∪ B] ≤ P [A] + P [B]
• P [A ∪ B] = P [A ∩ ¬B] + P [¬A ∩ B] + P [A ∩ B]
• P [A ∩ ¬B] = P [A] − P [A ∩ B]
Continuity of Probabilities
• A1 ⊂ A2 ⊂ . . . =⇒ limn→∞ P [An] = P [A] whereA =
∞
i=1 Ai
• A1 ⊃ A2 ⊃ . . . =⇒ limn→∞ P [An] = P [A] whereA =
∞
i=1 Ai
Independence ⊥⊥
A ⊥⊥ B ⇐⇒ P [A ∩ B] = P [A] P [B]
Conditional Probability
P [A | B] =
P [A ∩ B]
P [B]
P [B] > 0
Law of Total Probability
P [B] =
n
i=1
P [B|Ai] P [Ai] Ω =
n
i=1
Ai
Bayes’ Theorem
P [Ai | B] =
P [B | Ai] P [Ai]
n
j=1 P [B | Aj] P [Aj]
Ω =
n
i=1
Ai
Inclusion-Exclusion Principle
n
i=1
Ai =
n
r=1
(−1)r−1
i≤i1<···<ir≤n
r
j=1
Aij
3 Random Variables
Random Variable (RV)
X : Ω → R
Probability Mass Function (PMF)
fX(x) = P [X = x] = P [{ω ∈ Ω : X(ω) = x}]
Probability Density Function (PDF)
P [a ≤ X ≤ b] =
b
a
f(x) dx
Cumulative Distribution Function (CDF)
FX : R → [0, 1] FX(x) = P [X ≤ x]
1. Nondecreasing: x1 < x2 =⇒ F(x1) ≤ F(x2)
2. Normalized: limx→−∞ = 0 and limx→∞ = 1
3. Right-Continuous: limy↓x F(y) = F(x)
P [a ≤ Y ≤ b | X = x] =
b
a
fY |X(y | x)dy a ≤ b
fY |X(y | x) =
f(x, y)
fX(x)
Independence
1. P [X ≤ x, Y ≤ y] = P [X ≤ x] P [Y ≤ y]
2. fX,Y (x, y) = fX(x)fY (y)
6

3.1 Transformations
Transformation function
Z = ϕ(X)
Discrete
fZ(z) = P [ϕ(X) = z] = P [{x : ϕ(x) = z}] = P X ∈ ϕ−1
(z) =
x∈ϕ−1(z)
f(x)
Continuous
FZ(z) = P [ϕ(X) ≤ z] =
Az
f(x) dx with Az = {x : ϕ(x) ≤ z}
Special case if ϕ strictly monotone
fZ(z) = fX(ϕ−1
(z))
d
dz
ϕ−1
(z) = fX(x)
dx
dz
= fX(x)
1
|J|
The Rule of the Lazy Statistician
E [Z] = ϕ(x) dFX(x)
E [IA(x)] = IA(x) dFX(x) =
A
dFX(x) = P [X ∈ A]
Convolution
• Z := X + Y fZ(z) =
∞
−∞
fX,Y (x, z − x) dx
X,Y ≥0
=
z
0
fX,Y (x, z − x) dx
• Z := |X − Y | fZ(z) = 2
∞
0
fX,Y (x, z + x) dx
• Z :=
X
Y
fZ(z) =
∞
−∞
|x|fX,Y (x, xz) dx
⊥⊥
=
∞
−∞
xfx(x)fX(x)fY (xz) dx
4 Expectation
Definition and properties
• E [X] = µX = x dFX(x) =



x
xfX(x) X discrete
xfX(x) X continuous
• P [X = c] = 1 =⇒ E [c] = c
• E [cX] = c E [X]
• E [X + Y ] = E [X] + E [Y ]
• E [XY ] =
X,Y
xyfX,Y (x, y) dFX(x) dFY (y)
• E [ϕ(Y )] = ϕ(E [X]) (cf. Jensen inequality)
• P [X ≥ Y ] = 0 =⇒ E [X] ≥ E [Y ] ∧ P [X = Y ] = 1 =⇒ E [X] = E [Y ]
• E [X] =
∞
x=1
P [X ≥ x]
Sample mean
¯Xn =
1
n
n
i=1
Xi
Conditional expectation
• E [Y | X = x] = yf(y | x) dy
• E [X] = E [E [X | Y ]]
• E[ϕ(X, Y ) | X = x] =
∞
−∞
ϕ(x, y)fY |X(y | x) dx
• E [ϕ(Y, Z) | X = x] =
∞
−∞
ϕ(y, z)f(Y,Z)|X(y, z | x) dy dz
• E [Y + Z | X] = E [Y | X] + E [Z | X]
• E [ϕ(X)Y | X] = ϕ(X)E [Y | X]
• E[Y | X] = c =⇒ Cov [X, Y ] = 0
5 Variance
Definition and properties
• V [X] = σ2
X = E (X − E [X])2
= E X2
− E [X]
2
• V
n
i=1
Xi =
n
i=1
V [Xi] + 2
i=j
Cov [Xi, Yj]
• V
n
i=1
Xi =
n
i=1
V [Xi] iff Xi ⊥⊥ Xj
Standard deviation
sd[X] = V [X] = σX
Covariance
• Cov [X, Y ] = E [(X − E [X])(Y − E [Y ])] = E [XY ] − E [X] E [Y ]
• Cov [X, a] = 0
• Cov [X, X] = V [X]
• Cov [X, Y ] = Cov [Y, X]
• Cov [aX, bY ] = abCov [X, Y ]
7

• Cov [X + a, Y + b] = Cov [X, Y ]
• Cov


n
i=1
Xi,
m
j=1
Yj

 =
n
i=1
m
j=1
Cov [Xi, Yj]
Correlation
ρ [X, Y ] =
Cov [X, Y ]
V [X] V [Y ]
Independence
X ⊥⊥ Y =⇒ ρ [X, Y ] = 0 ⇐⇒ Cov [X, Y ] = 0 ⇐⇒ E [XY ] = E [X] E [Y ]
Sample variance
S2
=
1
n − 1
n
i=1
(Xi − ¯Xn)2
Conditional variance
• V [Y | X] = E (Y − E [Y | X])2
| X = E Y 2
| X − E [Y | X]
2
• V [Y ] = E [V [Y | X]] + V [E [Y | X]]
6 Inequalities
Cauchy-Schwarz
E [XY ]
2
≤ E X2
E Y 2
Markov
P [ϕ(X) ≥ t] ≤
E [ϕ(X)]
t
Chebyshev
P [|X − E [X]| ≥ t] ≤
V [X]
t2
Chernoff
P [X ≥ (1 + δ)µ] ≤
eδ
(1 + δ)1+δ
δ > −1
Jensen
E [ϕ(X)] ≥ ϕ(E [X]) ϕ convex
7 Distribution Relationships
Binomial
• Xi ∼ Bern (p) =⇒
n
i=1
Xi ∼ Bin (n, p)
• X ∼ Bin (n, p) , Y ∼ Bin (m, p) =⇒ X + Y ∼ Bin (n + m, p)
• limn→∞ Bin (n, p) = Po (np) (n large, p small)
• limn→∞ Bin (n, p) = N (np, np(1 − p)) (n large, p far from 0 and 1)
Negative Binomial
• X ∼ NBin (1, p) = Geo (p)
• X ∼ NBin (r, p) =
r
i=1 Geo (p)
• Xi ∼ NBin (ri, p) =⇒ Xi ∼ NBin ( ri, p)
• X ∼ NBin (r, p) . Y ∼ Bin (s + r, p) =⇒ P [X ≤ s] = P [Y ≥ r]
Poisson
• Xi ∼ Po (λi) ∧ Xi ⊥⊥ Xj =⇒
n
i=1
Xi ∼ Po
n
i=1
λi
• Xi ∼ Po (λi) ∧ Xi ⊥⊥ Xj =⇒ Xi
n
j=1
Xj ∼ Bin


n
j=1
Xj,
λi
n
j=1 λj


Exponential
• Xi ∼ Exp (β) ∧ Xi ⊥⊥ Xj =⇒
n
i=1
Xi ∼ Gamma (n, β)
• Memoryless property: P [X > x + y | X > y] = P [X > x]
Normal
• X ∼ N µ, σ2
=⇒ X−µ
σ ∼ N (0, 1)
• X ∼ N µ, σ2
∧ Z = aX + b =⇒ Z ∼ N aµ + b, a2
σ2
• X ∼ N µ1, σ2
1 ∧ Y ∼ N µ2, σ2
2 =⇒ X + Y ∼ N µ1 + µ2, σ2
1 + σ2
2
• Xi ∼ N µi, σ2
i =⇒ i Xi ∼ N i µi, i σ2
i
• P [a < X ≤ b] = Φ b−µ
σ − Φ a−µ
σ
• Φ(−x) = 1 − Φ(x) φ (x) = −xφ(x) φ (x) = (x2
− 1)φ(x)
• Upper quantile of N (0, 1): zα = Φ−1
(1 − α)
Gamma
• X ∼ Gamma (α, β) ⇐⇒ X/β ∼ Gamma (α, 1)
• Gamma (α, β) ∼
α
i=1 Exp (β)
• Xi ∼ Gamma (αi, β) ∧ Xi ⊥⊥ Xj =⇒ i Xi ∼ Gamma ( i αi, β)
•
Γ(α)
λα
=
∞
0
xα−1
e−λx
dx
Beta
•
1
B(α, β)
xα−1
(1 − x)β−1
=
Γ(α + β)
Γ(α)Γ(β)
xα−1
(1 − x)β−1
• E Xk
=
B(α + k, β)
B(α, β)
=
α + k − 1
α + β + k − 1
E Xk−1
• Beta (1, 1) ∼ Unif (0, 1)
8

8 Probability and Moment Generating Functions
• GX(t) = E tX
|t| < 1
• MX(t) = GX(et
) = E eXt
= E
∞
i=0
(Xt)i
i!
=
∞
i=0
E Xi
i!
· ti
• P [X = 0] = GX(0)
• P [X = 1] = GX(0)
• P [X = i] =
G
(i)
X (0)
i!
• E [X] = GX(1−
)
• E Xk
= M
(k)
X (0)
• E
X!
(X − k)!
= G
(k)
X (1−
)
• V [X] = GX(1−
) + GX(1−
) − (GX(1−
))
2
• GX(t) = GY (t) =⇒ X
d
= Y
9 Multivariate Distributions
9.1 Standard Bivariate Normal
Let X, Y ∼ N (0, 1) ∧ X ⊥⊥ Z where Y = ρX + 1 − ρ2Z
Joint density
f(x, y) =
1
2π 1 − ρ2
exp −
x2
+ y2
− 2ρxy
2(1 − ρ2)
Conditionals
(Y | X = x) ∼ N ρx, 1 − ρ2
and (X | Y = y) ∼ N ρy, 1 − ρ2
Independence
X ⊥⊥ Y ⇐⇒ ρ = 0
9.2 Bivariate Normal
Let X ∼ N µx, σ2
x and Y ∼ N µy, σ2
y .
f(x, y) =
1
2πσxσy 1 − ρ2
exp −
z
2(1 − ρ2)
z =
x − µx
σx
2
+
y − µy
σy
2
− 2ρ
x − µx
σx
y − µy
σy
Conditional mean and variance
E [X | Y ] = E [X] + ρ
σX
σY
(Y − E [Y ])
V [X | Y ] = σX 1 − ρ2
9.3 Multivariate Normal
Covariance matrix Σ (Precision matrix Σ−1
)
Σ =



V [X1] · · · Cov [X1, Xk]
...
...
...
Cov [Xk, X1] · · · V [Xk]



If X ∼ N (µ, Σ),
fX(x) = (2π)−n/2
|Σ|
−1/2
exp −
1
2
(x − µ)T
Σ−1
(x − µ)
Properties
• Z ∼ N (0, 1) ∧ X = µ + Σ1/2
Z =⇒ X ∼ N (µ, Σ)
• X ∼ N (µ, Σ) =⇒ Σ−1/2
(X − µ) ∼ N (0, 1)
• X ∼ N (µ, Σ) =⇒ AX ∼ N Aµ, AΣAT
• X ∼ N (µ, Σ) ∧ a = k =⇒ aT
X ∼ N aT
µ, aT
Σa
10 Convergence
Let {X1, X2, . . .} be a sequence of rv’s and let X be another rv. Let Fn denote
the cdf of Xn and let F denote the cdf of X.
Types of convergence
1. In distribution (weakly, in law): Xn
D
→ X
lim
n→∞
Fn(t) = F(t) ∀t where F continuous
2. In probability: Xn
P
→ X
(∀ε > 0) lim
n→∞
P [|Xn − X| > ε] = 0
3. Almost surely (strongly): Xn
as
→ X
P lim
n→∞
Xn = X = P ω ∈ Ω : lim
n→∞
Xn(ω) = X(ω) = 1
9

4. In quadratic mean (L2): Xn
qm
→ X
lim
n→∞
E (Xn − X)2
= 0
Relationships
• Xn
qm
→ X =⇒ Xn
P
→ X =⇒ Xn
D
→ X
• Xn
as
→ X =⇒ Xn
P
→ X
• Xn
D
→ X ∧ (∃c ∈ R) P [X = c] = 1 =⇒ Xn
P
→ X
• Xn
P
→ X ∧ Yn
P
→ Y =⇒ Xn + Yn
P
→ X + Y
• Xn
qm
→ X ∧ Yn
qm
→ Y =⇒ Xn + Yn
qm
→ X + Y
• Xn
P
→ X ∧ Yn
P
→ Y =⇒ XnYn
P
→ XY
• Xn
P
→ X =⇒ ϕ(Xn)
P
→ ϕ(X)
• Xn
D
→ X =⇒ ϕ(Xn)
D
→ ϕ(X)
• Xn
qm
→ b ⇐⇒ limn→∞ E [Xn] = b ∧ limn→∞ V [Xn] = 0
• X1, . . . , Xn iid ∧ E [X] = µ ∧ V [X] < ∞ ⇐⇒ ¯Xn
qm
→ µ
Slutzky’s Theorem
• Xn
D
→ X and Yn
P
→ c =⇒ Xn + Yn
D
→ X + c
• Xn
D
→ X and Yn
P
→ c =⇒ XnYn
D
→ cX
• In general: Xn
D
→ X and Yn
D
→ Y =⇒ Xn + Yn
D
→ X + Y
10.1 Law of Large Numbers (LLN)
Let {X1, . . . , Xn} be a sequence of iid rv’s, E [X1] = µ, and V [X1] < ∞.
Weak (WLLN)
¯Xn
P
→ µ n → ∞
Strong (WLLN)
¯Xn
as
→ µ n → ∞
10.2 Central Limit Theorem (CLT)
Let {X1, . . . , Xn} be a sequence of iid rv’s, E [X1] = µ, and V [X1] = σ2
.
Zn :=
¯Xn − µ
V ¯Xn
=
√
n( ¯Xn − µ)
σ
D
→ Z where Z ∼ N (0, 1)
lim
n→∞
P [Zn ≤ z] = Φ(z) z ∈ R
CLT notations
Zn ≈ N (0, 1)
¯Xn ≈ N µ,
σ2
n
¯Xn − µ ≈ N 0,
σ2
n
√
n( ¯Xn − µ) ≈ N 0, σ2
√
n( ¯Xn − µ)
n
≈ N (0, 1)
Continuity correction
P ¯Xn ≤ x ≈ Φ
x + 1
2 − µ
σ/
√
n
P ¯Xn ≥ x ≈ 1 − Φ
x − 1
2 − µ
σ/
√
n
Delta method
Yn ≈ N µ,
σ2
n
=⇒ ϕ(Yn) ≈ N ϕ(µ), (ϕ (µ))
2 σ2
n
11 Statistical Inference
Let X1, · · · , Xn
iid
∼ F if not otherwise noted.
11.1 Point Estimation
• Point estimator θn of θ is a rv: θn = g(X1, . . . , Xn)
• bias(θn) = E θn − θ
• Consistency: θn
P
→ θ
• Sampling distribution: F(θn)
• Standard error: se(θn) = V θn
• Mean squared error: mse = E (θn − θ)2
= bias(θn)2
+ V θn
• limn→∞ bias(θn) = 0 ∧ limn→∞ se(θn) = 0 =⇒ θn is consistent
• Asymptotic normality:
θn − θ
se
D
→ N (0, 1)
• Slutzky’s Theorem often lets us replace se(θn) by some (weakly) consis-
tent estimator σn.
10

11.2 Normal-Based Confidence Interval
Suppose θn ≈ N θ, se2
. Let zα/2 = Φ−1
(1 − (α/2)), i.e., P Z > zα/2 = α/2
and P −zα/2 < Z < zα/2 = 1 − α where Z ∼ N (0, 1). Then
Cn = θn ± zα/2se
11.3 Empirical distribution
Empirical Distribution Function (ECDF)
Fn(x) =
n
i=1 I(Xi ≤ x)
n
I(Xi ≤ x) =
1 Xi ≤ x
0 Xi > x
Properties (for any fixed x)
• E Fn = F(x)
• V Fn =
F(x)(1 − F(x))
n
• mse =
F(x)(1 − F(x))
n
D
→ 0
• Fn
P
→ F(x)
Dvoretzky-Kiefer-Wolfowitz (DKW) inequality (X1, . . . , Xn ∼ F)
P sup
x
F(x) − Fn(x) > ε = 2e−2nε2
Nonparametric 1 − α confidence band for F
L(x) = max{Fn − n, 0}
U(x) = min{Fn + n, 1}
=
1
2n
log
2
α
P [L(x) ≤ F(x) ≤ U(x) ∀x] ≥ 1 − α
11.4 Statistical Functionals
• Statistical functional: T(F)
• Plug-in estimator of θ = (F): θn = T(Fn)
• Linear functional: T(F) = ϕ(x) dFX(x)
• Plug-in estimator for linear functional:
T(Fn) = ϕ(x) dFn(x) =
1
n
n
i=1
ϕ(Xi)
• Often: T(Fn) ≈ N T(F), se2
=⇒ T(Fn) ± zα/2se
• pth
quantile: F−1
(p) = inf{x : F(x) ≥ p}
• µ = ¯Xn
• σ2
=
1
n − 1
n
i=1
(Xi − ¯Xn)2
• κ =
1
n
n
i=1(Xi − µ)3
σ3j
• ρ =
n
i=1(Xi − ¯Xn)(Yi − ¯Yn)
n
i=1(Xi − ¯Xn)2 n
i=1(Yi − ¯Yn)
12 Parametric Inference
Let F = f(x; θ) : θ ∈ Θ be a parametric model with parameter space Θ ⊂ Rk
and parameter θ = (θ1, . . . , θk).
12.1 Method of Moments
jth
moment
αj(θ) = E Xj
= xj
dFX(x)
jth
sample moment
αj =
1
n
n
i=1
Xj
i
Method of moments estimator (MoM)
α1(θ) = α1
α2(θ) = α2
... =
...
αk(θ) = αk
11

Properties of the MoM estimator
• θn exists with probability tending to 1
P
→ θ
√
n(θ − θ)
D
→ N (0, Σ)
where Σ = gE Y Y T
gT
, Y = (X, X2
, . . . , Xk
)T
,
g = (g1, . . . , gk) and gj = ∂
∂θ α−1
j (θ)
12.2 Maximum Likelihood
Likelihood: Ln : Θ → [0, ∞)
Ln(θ) =
n
i=1
f(Xi; θ)
Log-likelihood
n(θ) = log Ln(θ) =
n
i=1
log f(Xi; θ)
Maximum likelihood estimator (mle)
Ln(θn) = sup
θ
Ln(θ)
Score function
s(X; θ) =
∂
∂θ
log f(X; θ)
Fisher information
I(θ) = Vθ [s(X; θ)]
In(θ) = nI(θ)
Fisher information (exponential family)
I(θ) = Eθ −
∂
∂θ
s(X; θ)
Observed Fisher information
Iobs
n (θ) = −
∂2
∂θ2
n
i=1
log f(Xi; θ)
Properties of the mle
P
→ θ
• Equivariance: θn is the mle =⇒ ϕ(θn) ist the mle of ϕ(θ)
1. se ≈ 1/In(θ)
(θn − θ)
se
D
→ N (0, 1)
2. se ≈ 1/In(θn)
(θn − θ)
se
D
→ N (0, 1)
• Asymptotic optimality (or efficiency), i.e., smallest variance for large sam-
ples. If θn is any other estimator, the asymptotic relative efficiency is
are(θn, θn) =
V θn
V θn
≤ 1
• Approximately the Bayes estimator
12.2.1 Delta Method
If τ = ϕ(θ) where ϕ is differentiable and ϕ (θ) = 0:
(τn − τ)
se(τ)
D
→ N (0, 1)
where τ = ϕ(θ) is the mle of τ and
se = ϕ (θ) se(θn)
12.3 Multiparameter Models
Let θ = (θ1, . . . , θk) and θ = (θ1, . . . , θk) be the mle.
Hjj =
∂2
n
∂θ2
Hjk =
∂2
n
∂θj∂θk
Fisher information matrix
In(θ) = −



Eθ [H11] · · · Eθ [H1k]
...
...
...
Eθ [Hk1] · · · Eθ [Hkk]



Under appropriate regularity conditions
(θ − θ) ≈ N (0, Jn)
12

with Jn(θ) = I−1
n . Further, if θj is the jth
component of θ, then
(θj − θj)
sej
D
→ N (0, 1)
where se2
j = Jn(j, j) and Cov θj, θk = Jn(j, k)
12.3.1 Multiparameter delta method
Let τ = ϕ(θ1, . . . , θk) and let the gradient of ϕ be
ϕ =






∂ϕ
∂θ1
...
∂ϕ
∂θk






Suppose ϕ θ=θ
= 0 and τ = ϕ(θ). Then,
(τ − τ)
se(τ)
D
→ N (0, 1)
where
se(τ) = ϕ
T
Jn ϕ
and Jn = Jn(θ) and ϕ = ϕ θ=θ
.
12.4 Parametric Bootstrap
Sample from f(x; θn) instead of from Fn, where θn could be the mle or method
of moments estimator.
13 Hypothesis Testing
H0 : θ ∈ Θ0 versus H1 : θ ∈ Θ1
Deﬁnitions
• Null hypothesis H0
• Alternative hypothesis H1
• Simple hypothesis θ = θ0
• Composite hypothesis θ > θ0 or θ < θ0
• Two-sided test: H0 : θ = θ0 versus H1 : θ = θ0
• One-sided test: H0 : θ ≤ θ0 versus H1 : θ > θ0
• Critical value c
• Test statistic T
• Rejection region R = {x : T(x) > c}
• Power function β(θ) = P [X ∈ R]
• Power of a test: 1 − P [Type II error] = 1 − β = inf
θ∈Θ1
β(θ)
• Test size: α = P [Type I error] = sup
θ∈Θ0
β(θ)
Retain H0 Reject H0
H0 true
√
Type I Error (α)
H1 true Type II Error (β)
√
(power)
p-value
• p-value = supθ∈Θ0
Pθ [T(X) ≥ T(x)] = inf α : T(x) ∈ Rα
• p-value = supθ∈Θ0
Pθ [T(X ) ≥ T(X)]
1−Fθ(T (X)) since T (X )∼Fθ
= inf α : T(X) ∈ Rα
p-value evidence
< 0.01 very strong evidence against H0
0.01 − 0.05 strong evidence against H0
0.05 − 0.1 weak evidence against H0
> 0.1 little or no evidence against H0
Wald test
• Two-sided test
• Reject H0 when |W| > zα/2 where W =
θ − θ0
se
• P |W| > zα/2 → α
• p-value = Pθ0
[|W| > |w|] ≈ P [|Z| > |w|] = 2Φ(−|w|)
Likelihood ratio test (LRT)
• T(X) =
supθ∈Θ Ln(θ)
supθ∈Θ0
Ln(θ)
=
Ln(θn)
Ln(θn,0) 13

• λ(X) = 2 log T(X)
D
→ χ2
r−q where
k
i=1
Z2
i ∼ χ2
k and Z1, . . . , Zk
iid
∼ N (0, 1)
• p-value = Pθ0 [λ(X) > λ(x)] ≈ P χ2
r−q > λ(x)
Multinomial LRT
• mle: pn =
X1
n
, . . . ,
Xk
n
• T(X) =
Ln(pn)
Ln(p0)
=
k
j=1
pj
p0j
Xj
• λ(X) = 2
k
j=1
Xj log
pj
p0j
D
→ χ2
k−1
• The approximate size α LRT rejects H0 when λ(X) ≥ χ2
k−1,α
Pearson Chi-square Test
• T =
k
j=1
(Xj − E [Xj])2
E [Xj]
where E [Xj] = np0j under H0
• T
D
→ χ2
k−1
• p-value = P χ2
k−1 > T(x)
• Faster
D
→ X2
k−1 than LRT, hence preferable for small n
Independence testing
• I rows, J columns, X multinomial sample of size n = I ∗ J
• mles unconstrained: pij =
Xij
n
• mles under H0: p0ij = pi·p·j = Xi·
n
X·j
n
• LRT: λ = 2
I
i=1
J
j=1 Xij log
nXij
Xi·X·j
• PearsonChiSq: T =
I
i=1
J
j=1
(Xij −E[Xij ])2
E[Xij ]
• LRT and Pearson
D
→ χ2
kν, where ν = (I − 1)(J − 1)
14 Bayesian Inference
Bayes’ Theorem
f(θ | x) =
f(x | θ)f(θ)
f(xn)
=
f(x | θ)f(θ)
f(x | θ)f(θ) dθ
∝ Ln(θ)f(θ)
Deﬁnitions
• Xn
= (X1, . . . , Xn)
• xn
= (x1, . . . , xn)
• Prior density f(θ)
• Likelihood f(xn
| θ): joint density of the data
In particular, Xn
iid =⇒ f(xn
| θ) =
n
i=1
f(xi | θ) = Ln(θ)
• Posterior density f(θ | xn
)
• Normalizing constant cn = f(xn
) = f(x | θ)f(θ) dθ
• Kernel: part of a density that depends on θ
• Posterior mean ¯θn = θf(θ | xn
) dθ =
θLn(θ)f(θ)
Ln(θ)f(θ) dθ
14.1 Credible Intervals
Posterior interval
P [θ ∈ (a, b) | xn
] =
b
a
f(θ | xn
) dθ = 1 − α
Equal-tail credible interval
a
−∞
f(θ | xn
) dθ =
∞
b
f(θ | xn
) dθ = α/2
Highest posterior density (HPD) region Rn
1. P [θ ∈ Rn] = 1 − α
2. Rn = {θ : f(θ | xn
) > k} for some k
Rn is unimodal =⇒ Rn is an interval
14.2 Function of parameters
Let τ = ϕ(θ) and A = {θ : ϕ(θ) ≤ τ}.
Posterior CDF for τ
H(r | xn
) = P [ϕ(θ) ≤ τ | xn
] =
A
f(θ | xn
) dθ
Posterior density
h(τ | xn
) = H (τ | xn
)
Bayesian delta method
τ | Xn
≈ N ϕ(θ), se ϕ (θ)
14

14.3 Priors
Choice
• Subjective bayesianism.
• Objective bayesianism.
• Robust bayesianism.
Types
• Flat: f(θ) ∝ constant
• Proper:
∞
−∞
f(θ) dθ = 1
• Improper:
∞
−∞
f(θ) dθ = ∞
• Jeffrey’s prior (transformation-invariant):
f(θ) ∝ I(θ) f(θ) ∝ det(I(θ))
• Conjugate: f(θ) and f(θ | xn
) belong to the same parametric family
14.3.1 Conjugate Priors
Discrete likelihood
Likelihood Conjugate prior Posterior hyperparameters
Bern (p) Beta (α, β) α +
n
i=1
xi, β + n −
n
i=1
xi
Bin (p) Beta (α, β) α +
n
i=1
xi, β +
n
i=1
Ni −
n
i=1
xi
NBin (p) Beta (α, β) α + rn, β +
n
i=1
xi
Po (λ) Gamma (α, β) α +
n
i=1
xi, β + n
Multinomial(p) Dir (α) α +
n
i=1
x(i)
Geo (p) Beta (α, β) α + n, β +
n
i=1
xi
Continuous likelihood (subscript c denotes constant)
Likelihood Conjugate prior Posterior hyperparameters
Unif (0, θ) Pareto(xm, k) max x(n), xm , k + n
Exp (λ) Gamma (α, β) α + n, β +
n
i=1
xi
N µ, σ2
c N µ0, σ2
0
µ0
σ2
0
+
n
i=1 xi
σ2
c
/
1
σ2
0
+
n
σ2
c
,
1
σ2
0
+
n
σ2
c
−1
N µc, σ2
Scaled Inverse Chi-
square(ν, σ2
0)
ν + n,
νσ2
0 +
n
i=1(xi − µ)2
ν + n
N µ, σ2
Normal-
scaled Inverse
Gamma(λ, ν, α, β)
νλ + n¯x
ν + n
, ν + n, α +
n
2
,
β +
1
2
n
i=1
(xi − ¯x)2
+
γ(¯x − λ)2
2(n + γ)
MVN(µ, Σc) MVN(µ0, Σ0) Σ−1
0 + nΣ−1
c
−1
Σ−1
0 µ0 + nΣ−1
¯x ,
Σ−1
0 + nΣ−1
c
−1
MVN(µc, Σ) Inverse-
Wishart(κ, Ψ)
n + κ, Ψ +
n
i=1
(xi − µc)(xi − µc)T
Pareto(xmc
, k) Gamma (α, β) α + n, β +
n
i=1
log
xi
xmc
Pareto(xm, kc) Pareto(x0, k0) x0, k0 − kn where k0 > kn
Gamma (αc, β) Gamma (α0, β0) α0 + nαc, β0 +
n
i=1
xi
14.4 Bayesian Testing
If H0 : θ ∈ Θ0:
Prior probability P [H0] =
Θ0
f(θ) dθ
Posterior probability P [H0 | xn
] =
Θ0
f(θ | xn
) dθ
Let H0, . . . , HK−1 be K hypotheses. Suppose θ ∼ f(θ | Hk),
P [Hk | xn
] =
f(xn
| Hk)P [Hk]
K
k=1 f(xn | Hk)P [Hk]
,
15

Marginal likelihood
f(xn
| Hi) =
Θ
f(xn
| θ, Hi)f(θ | Hi) dθ
Posterior odds (of Hi relative to Hj)
P [Hi | xn
]
P [Hj | xn]
=
f(xn
| Hi)
f(xn | Hj)
Bayes Factor BFij
×
P [Hi]
P [Hj]
prior odds
Bayes factor
log10 BF10 BF10 evidence
0 − 0.5 1 − 1.5 Weak
0.5 − 1 1.5 − 10 Moderate
1 − 2 10 − 100 Strong
> 2 > 100 Decisive
p∗
=
p
1−p BF10
1 + p
1−p BF10
where p = P [H1] and p∗
= P [H1 | xn
]
15 Exponential Family
Scalar parameter
fX(x | θ) = h(x) exp {η(θ)T(x) − A(θ)}
= h(x)g(θ) exp {η(θ)T(x)}
Vector parameter
fX(x | θ) = h(x) exp
s
i=1
ηi(θ)Ti(x) − A(θ)
= h(x) exp {η(θ) · T(x) − A(θ)}
= h(x)g(θ) exp {η(θ) · T(x)}
Natural form
fX(x | η) = h(x) exp {η · T(x) − A(η)}
= h(x)g(η) exp {η · T(x)}
= h(x)g(η) exp ηT
T(x)
16 Sampling Methods
16.1 The Bootstrap
Let Tn = g(X1, . . . , Xn) be a statistic.
1. Estimate VF [Tn] with VFn
[Tn].
2. Approximate VFn
[Tn] using simulation:
(a) Repeat the following B times to get T∗
n,1, . . . , T∗
n,B, an iid sample from
the sampling distribution implied by Fn
i. Sample uniformly X∗
1 , . . . , X∗
n ∼ Fn.
ii. Compute T∗
n = g(X∗
1 , . . . , X∗
n).
(b) Then
vboot = VFn
=
1
B
B
b=1
T∗
n,b −
1
B
B
r=1
T∗
n,r
2
16.1.1 Bootstrap Confidence Intervals
Normal-based interval
Tn ± zα/2seboot
Pivotal interval
1. Location parameter θ = T(F)
2. Pivot Rn = θn − θ
3. Let H(r) = P [Rn ≤ r] be the cdf of Rn
4. Let R∗
n,b = θ∗
n,b − θn. Approximate H using bootstrap:
H(r) =
1
B
B
b=1
I(R∗
n,b ≤ r)
5. θ∗
β = β sample quantile of (θ∗
n,1, . . . , θ∗
n,B)
6. r∗
β = β sample quantile of (R∗
n,1, . . . , R∗
n,B), i.e., r∗
β = θ∗
β − θn
7. Approximate 1 − α confidence interval Cn = â,ˆb where
â = θn − H−1
1 −
α
2
= θn − r∗
1−α/2 = 2θn − θ∗
1−α/2
ˆb = θn − H−1 α
2
= θn − r∗
α/2 = 2θn − θ∗
α/2
Percentile interval
Cn = θ∗
α/2, θ∗
1−α/2
16

16.2 Rejection Sampling
Setup
• We can easily sample from g(θ)
• We want to sample from h(θ), but it is difficult
• We know h(θ) up to a proportional constant: h(θ) =
k(θ)
k(θ) dθ
• Envelope condition: we can find M > 0 such that k(θ) ≤ Mg(θ) ∀θ
Algorithm
1. Draw θcand
∼ g(θ)
2. Generate u ∼ Unif (0, 1)
3. Accept θcand
if u ≤
k(θcand
)
Mg(θcand)
4. Repeat until B values of θcand
have been accepted
Example
• We can easily sample from the prior g(θ) = f(θ)
• Target is the posterior h(θ) ∝ k(θ) = f(xn
| θ)f(θ)
• Envelope condition: f(xn
| θ) ≤ f(xn
| θn) = Ln(θn) ≡ M
• Algorithm
1. Draw θcand
∼ f(θ)
2. Generate u ∼ Unif (0, 1)
3. Accept θcand
if u ≤
Ln(θcand
)
Ln(θn)
16.3 Importance Sampling
Sample from an importance function g rather than target density h.
Algorithm to obtain an approximation to E [q(θ) | xn
]:
1. Sample from the prior θ1, . . . , θn
iid
∼ f(θ)
2. wi =
Ln(θi)
B
i=1 Ln(θi)
∀i = 1, . . . , B
3. E [q(θ) | xn
] ≈
B
i=1 q(θi)wi
17 Decision Theory
Definitions
• Unknown quantity affecting our decision: θ ∈ Θ
• Decision rule: synonymous for an estimator θ
• Action a ∈ A: possible value of the decision rule. In the estimation
context, the action is just an estimate of θ, θ(x).
• Loss function L: consequences of taking action a when true state is θ or
discrepancy between θ and θ, L : Θ × A → [−k, ∞).
Loss functions
• Squared error loss: L(θ, a) = (θ − a)2
• Linear loss: L(θ, a) =
K1(θ − a) a − θ < 0
K2(a − θ) a − θ ≥ 0
• Absolute error loss: L(θ, a) = |θ − a| (linear loss with K1 = K2)
• Lp loss: L(θ, a) = |θ − a|p
• Zero-one loss: L(θ, a) =
0 a = θ
1 a = θ
17.1 Risk
Posterior risk
r(θ | x) = L(θ, θ(x))f(θ | x) dθ = Eθ|X L(θ, θ(x))
(Frequentist) risk
R(θ, θ) = L(θ, θ(x))f(x | θ) dx = EX|θ L(θ, θ(X))
Bayes risk
r(f, θ) = L(θ, θ(x))f(x, θ) dx dθ = Eθ,X L(θ, θ(X))
r(f, θ) = Eθ EX|θ L(θ, θ(X) = Eθ R(θ, θ)
r(f, θ) = EX Eθ|X L(θ, θ(X) = EX r(θ | X)
17.2 Admissibility
• θ dominates θ if
∀θ : R(θ, θ ) ≤ R(θ, θ)
∃θ : R(θ, θ ) < R(θ, θ)
• θ is inadmissible if there is at least one other estimator θ that dominates
it. Otherwise it is called admissible.
17

17.3 Bayes Rule
Bayes rule (or Bayes estimator)
• r(f, θ) = infθ r(f, θ)
• θ(x) = inf r(θ | x) ∀x =⇒ r(f, θ) = r(θ | x)f(x) dx
Theorems
• Squared error loss: posterior mean
• Absolute error loss: posterior median
• Zero-one loss: posterior mode
17.4 Minimax Rules
Maximum risk
¯R(θ) = sup
θ
R(θ, θ) ¯R(a) = sup
θ
R(θ, a)
Minimax rule
sup
θ
R(θ, θ) = inf
θ
¯R(θ) = inf
θ
sup
θ
R(θ, θ)
θ = Bayes rule ∧ ∃c : R(θ, θ) = c
Least favorable prior
θf
= Bayes rule ∧ R(θ, θf
) ≤ r(f, θf
) ∀θ
18 Linear Regression
Definitions
• Response variable Y
• Covariate X (aka predictor variable or feature)
18.1 Simple Linear Regression
Model
Yi = β0 + β1Xi + i E [ i | Xi] = 0, V [ i | Xi] = σ2
Fitted line
r(x) = β0 + β1x
Predicted (fitted) values
Yi = r(Xi)
Residuals
î = Yi − Yi = Yi − β0 + β1Xi
Residual sums of squares (rss)
rss(β0, β1) =
n
i=1
ˆ2
i
Least square estimates
βT
= (β0, β1)T
: min
β0,β1
rss
β0 = ¯Yn − β1
¯Xn
β1 =
n
i=1(Xi − ¯Xn)(Yi − ¯Yn)
n
i=1(Xi − ¯Xn)2
=
n
i=1 XiYi − n ¯XY
n
i=1 X2
i − nX2
E β | Xn
=
β0
β1
V β | Xn
=
σ2
nsX
n−1 n
i=1 X2
i −Xn
−Xn 1
se(β0) =
σ
sX
√
n
n
i=1 X2
i
n
se(β1) =
σ
sX
√
n
where s2
X = n−1 n
i=1(Xi − Xn)2
and σ2
= 1
n−2
n
i=1 ˆ2
i (unbiased estimate).
Further properties:
• Consistency: β0
P
→ β0 and β1
P
→ β1
β0 − β0
se(β0)
D
→ N (0, 1) and
β1 − β1
se(β1)
D
→ N (0, 1)
• Approximate 1 − α confidence intervals for β0 and β1:
β0 ± zα/2se(β0) and β1 ± zα/2se(β1)
• Wald test for H0 : β1 = 0 vs. H1 : β1 = 0: reject H0 if |W| > zα/2 where
W = β1/se(β1).
R2
R2
=
n
i=1(Yi − Y )2
n
i=1(Yi − Y )2
= 1 −
n
i=1 ˆ2
i
n
i=1(Yi − Y )2
= 1 −
rss
tss
18

Likelihood
L =
n
i=1
f(Xi, Yi) =
n
i=1
fX(Xi) ×
n
i=1
fY |X(Yi | Xi) = L1 × L2
L1 =
n
i=1
fX(Xi)
L2 =
n
i=1
fY |X(Yi | Xi) ∝ σ−n
exp −
1
2σ2
i
Yi − (β0 − β1Xi)
2
Under the assumption of Normality, the least squares estimator is also the mle
σ2
=
1
n
n
i=1
ˆ2
i
18.2 Prediction
Observe X = x∗ of the covariate and want to predict their outcome Y∗.
Y∗ = β0 + β1x∗
V Y∗ = V β0 + x2
∗V β1 + 2x∗Cov β0, β1
Prediction interval
ξ2
n = σ2
n
i=1(Xi − X∗)2
n i(Xi − ¯X)2j
+ 1
Y∗ ± zα/2ξn
18.3 Multiple Regression
Y = Xβ +
where
X =



X11 · · · X1k
...
...
...
Xn1 · · · Xnk


 β =



β1
...
βk


 =



1
...
n



Likelihood
L(µ, Σ) = (2πσ2
)−n/2
exp −
1
2σ2
rss
rss = (y − Xβ)T
(y − Xβ) = Y − Xβ 2
=
N
i=1
(Yi − xT
i β)2
If the (k × k) matrix XT
X is invertible,
β = (XT
X)−1
XT
Y
V β | Xn
= σ2
(XT
X)−1
β ≈ N β, σ2
(XT
X)−1
Estimate regression function
r(x) =
k
j=1
βjxj
Unbiased estimate for σ2
σ2
=
1
n − k
n
i=1
ˆ2
i ˆ = Xβ − Y
mle
µ = ¯X σ2
=
n − k
n
σ2
1 − α Confidence interval
βj ± zα/2se(βj)
18.4 Model Selection
Consider predicting a new observation Y ∗
for covariates X∗
and let S ⊂ J
denote a subset of the covariates in the model, where |S| = k and |J| = n.
Issues
• Underfitting: too few covariates yields high bias
• Overfitting: too many covariates yields high variance
Procedure
1. Assign a score to each model
2. Search through all models to find the one with the highest score
Hypothesis testing
H0 : βj = 0 vs. H1 : βj = 0 ∀j ∈ J
Mean squared prediction error (mspe)
mspe = E (Y (S) − Y ∗
)2
Prediction risk
R(S) =
n
i=1
mspei =
n
i=1
E (Yi(S) − Y ∗
i )2
Training error
Rtr(S) =
n
i=1
(Yi(S) − Yi)2
19

R2
R2
(S) = 1 −
rss(S)
tss
= 1 −
Rtr(S)
tss
= 1 −
n
i=1(Yi(S) − Y )2
n
i=1(Yi − Y )2
The training error is a downward-biased estimate of the prediction risk.
E Rtr(S) < R(S)
bias(Rtr(S)) = E Rtr(S) − R(S) = −2
n
i=1
Cov Yi, Yi
Adjusted R2
R2
(S) = 1 −
n − 1
n − k
rss
tss
Mallow’s Cp statistic
R(S) = Rtr(S) + 2kσ2
= lack of fit + complexity penalty
Akaike Information Criterion (AIC)
AIC(S) = n(βS, σ2
S) − k
Bayesian Information Criterion (BIC)
BIC(S) = n(βS, σ2
S) −
k
2
log n
Validation and training
RV (S) =
m
i=1
(Y ∗
i (S) − Y ∗
i )2
m = |{validation data}|, often
n
4
or
n
2
Leave-one-out cross-validation
RCV (S) =
n
i=1
(Yi − Y(i))2
=
n
i=1
Yi − Yi(S)
1 − Uii(S)
2
U(S) = XS(XT
S XS)−1
XS (“hat matrix”)
19 Non-parametric Function Estimation
19.1 Density Estimation
Estimate f(x), where f(x) = P [X ∈ A] = A
f(x) dx.
Integrated square error (ise)
L(f, fn) = f(x) − fn(x)
2
dx = J(h) + f2
(x) dx
Frequentist risk
R(f, fn) = E L(f, fn) = b2
(x) dx + v(x) dx
b(x) = E fn(x) − f(x)
v(x) = V fn(x)
19.1.1 Histograms
Definitions
• Number of bins m
• Binwidth h = 1
m
• Bin Bj has νj observations
• Define pj = νj/n and pj = Bj
f(u) du
Histogram estimator
fn(x) =
m
j=1
pj
h
I(x ∈ Bj)
E fn(x) =
pj
h
V fn(x) =
pj(1 − pj)
nh2
R(fn, f) ≈
h2
12
(f (u))
2
du +
1
nh
h∗
=
1
n1/3
6
(f (u))
2 du
1/3
R∗
(fn, f) ≈
C
n2/3
C =
3
4
2/3
(f (u))
2
du
1/3
Cross-validation estimate of E [J(h)]
JCV (h) = f2
n(x) dx −
2
n
n
i=1
f(−i)(Xi) =
2
(n − 1)h
−
n + 1
(n − 1)h
m
j=1
p2
j
20

19.1.2 Kernel Density Estimator (KDE)
Kernel K
• K(x) ≥ 0
• K(x) dx = 1
• xK(x) dx = 0
• x2
K(x) dx ≡ σ2
K > 0
KDE
fn(x) =
1
n
n
i=1
1
h
K
x − Xi
h
R(f, fn) ≈
1
4
(hσK)4
(f (x))2
dx +
1
nh
K2
(x) dx
h∗
=
c
−2/5
1 c
−1/5
2 c
−1/5
3
n1/5
c1 = σ2
K, c2 = K2
(x) dx, c3 = (f (x))2
dx
R∗
(f, fn) =
c4
n4/5
c4 =
5
4
(σ2
K)2/5
K2
(x) dx
4/5
C(K)
(f )2
dx
1/5
Epanechnikov Kernel
K(x) =
3
4
√
5(1−x2/5)
|x| <
√
5
0 otherwise
JCV (h) = f2
n(x) dx −
2
n
n
i=1
f(−i)(Xi) ≈
1
hn2
n
i=1
n
j=1
K∗ Xi − Xj
h
+
2
nh
K(0)
K∗
(x) = K(2)
(x) − 2K(x) K(2)
(x) = K(x − y)K(y) dy
19.2 Non-parametric Regression
Estimate f(x) where f(x) = E [Y | X = x]. Consider pairs of points
(x1, Y1), . . . , (xn, Yn) related by
Yi = r(xi) + i
E [ i] = 0
V [ i] = σ2
k-nearest Neighbor Estimator
r(x) =
1
k
i:xi∈Nk(x)
Yi where Nk(x) = {k values of x1, . . . , xn closest to x}
Nadaraya-Watson Kernel Estimator
r(x) =
n
i=1
wi(x)Yi
wi(x) =
K x−xi
h
n
j=1 K
x−xj
h
∈ [0, 1]
R(rn, r) ≈
h4
4
x2
K2
(x) dx
4
r (x) + 2r (x)
f (x)
f(x)
2
dx
+
σ2
K2
(x) dx
nhf(x)
dx
h∗
≈
c1
n1/5
R∗
(rn, r) ≈
c2
n4/5
JCV (h) =
n
i=1
(Yi − r(−i)(xi))2
=
n
i=1
(Yi − r(xi))2
1 − K(0)
n
j=1 K
x−xj
h
2
19.3 Smoothing Using Orthogonal Functions
Approximation
r(x) =
∞
j=1
βjφj(x) ≈
J
i=1
βjφj(x)
Multivariate regression
Y = Φβ + η
where ηi = i and Φ =



φ0(x1) · · · φJ (x1)
...
...
...
φ0(xn) · · · φJ (xn)



Least squares estimator
β = (ΦT
Φ)−1
ΦT
Y
≈
1
n
ΦT
Y (for equally spaced observations only)
21

RCV (J) =
n
i=1

Yi −
J
j=1
φj(xi)βj,(−i)


2
20 Stochastic Processes
Stochastic Process
{Xt : t ∈ T} T =
{0, ±1, . . . } = Z discrete
[0, ∞) continuous
• Notations Xt, X(t)
• State space X
• Index set T
20.1 Markov Chains
Markov chain
P [Xn = x | X0, . . . , Xn−1] = P [Xn = x | Xn−1] ∀n ∈ T, x ∈ X
Transition probabilities
pij ≡ P [Xn+1 = j | Xn = i]
pij(n) ≡ P [Xm+n = j | Xm = i] n-step
Transition matrix P (n-step: Pn)
• (i, j) element is pij
• pij > 0
• i pij = 1
Chapman-Kolmogorov
pij(m + n) =
k
pij(m)pkj(n)
Pm+n = PmPn
Pn = P × · · · × P = Pn
Marginal probability
µn = (µn(1), . . . , µn(N)) where µi(i) = P [Xn = i]
µ0 initial distribution
µn = µ0Pn
20.2 Poisson Processes
Poisson process
• {Xt : t ∈ [0, ∞)} = number of events up to and including time t
• X0 = 0
• Independent increments:
∀t0 < · · · < tn : Xt1 − Xt0 ⊥⊥ · · · ⊥⊥ Xtn − Xtn−1
• Intensity function λ(t)
– P [Xt+h − Xt = 1] = λ(t)h + o(h)
– P [Xt+h − Xt = 2] = o(h)
• Xs+t − Xs ∼ Po (m(s + t) − m(s)) where m(t) =
t
0
λ(s) ds
Homogeneous Poisson process
λ(t) ≡ λ =⇒ Xt ∼ Po (λt) λ > 0
Waiting times
Wt := time at which Xt occurs
Wt ∼ Gamma t,
1
λ
Interarrival times
St = Wt+1 − Wt
St ∼ Exp
1
λ
tWt−1 Wt
St
22

21 Time Series
Mean function
µxt
= E [xt] =
∞
−∞
xft(x) dx
Autocovariance function
γx(s, t) = E [(xs − µs)(xt − µt)] = E [xsxt] − µsµt
γx(t, t) = E (xt − µt)2
= V [xt]
Autocorrelation function (ACF)
ρ(s, t) =
Cov [xs, xt]
V [xs] V [xt]
=
γ(s, t)
γ(s, s)γ(t, t)
Cross-covariance function (CCV)
γxy(s, t) = E [(xs − µxs )(yt − µyt )]
Cross-correlation function (CCF)
ρxy(s, t) =
γxy(s, t)
γx(s, s)γy(t, t)
Backshift operator
Bk
(xt) = xt−k
Diﬀerence operator
d
= (1 − B)d
White noise
• wt ∼ wn(0, σ2
w)
• Gaussian: wt
iid
∼ N 0, σ2
w
• E [wt] = 0 t ∈ T
• V [wt] = σ2
t ∈ T
• γw(s, t) = 0 s = t ∧ s, t ∈ T
Random walk
• Drift δ
• xt = δt +
t
j=1 wj
• E [xt] = δt
Symmetric moving average
mt =
k
j=−k
ajxt−j where aj = a−j ≥ 0 and
k
j=−k
aj = 1
21.1 Stationary Time Series
Strictly stationary
P [xt1 ≤ c1, . . . , xtk
≤ ck] = P [xt1+h ≤ c1, . . . , xtk+h ≤ ck]
∀k ∈ N, tk, ck, h ∈ Z
Weakly stationary
• E x2
t < ∞ ∀t ∈ Z
• E x2
t = m ∀t ∈ Z
• γx(s, t) = γx(s + r, t + r) ∀r, s, t ∈ Z
Autocovariance function
• γ(h) = E [(xt+h − µ)(xt − µ)] ∀h ∈ Z
• γ(0) = E (xt − µ)2
• γ(0) ≥ 0
• γ(0) ≥ |γ(h)|
• γ(h) = γ(−h)
Autocorrelation function (ACF)
ρx(h) =
Cov [xt+h, xt]
V [xt+h] V [xt]
=
γ(t + h, t)
γ(t + h, t + h)γ(t, t)
=
γ(h)
γ(0)
Jointly stationary time series
γxy(h) = E [(xt+h − µx)(yt − µy)]
ρxy(h) =
γxy(h)
γx(0)γy(h)
Linear process
xt = µ +
∞
j=−∞
ψjwt−j where
∞
j=−∞
|ψj| < ∞
γ(h) = σ2
w
∞
j=−∞
ψj+hψj
23

21.2 Estimation of Correlation
Sample mean
¯x =
1
n
n
t=1
xt
Sample variance
V [¯x] =
1
n
n
h=−n
1 −
|h|
n
γx(h)
Sample autocovariance function
γ(h) =
1
n
n−h
t=1
(xt+h − ¯x)(xt − ¯x)
Sample autocorrelation function
ρ(h) =
γ(h)
γ(0)
Sample cross-variance function
γxy(h) =
1
n
n−h
t=1
(xt+h − ¯x)(yt − y)
Sample cross-correlation function
ρxy(h) =
γxy(h)
γx(0)γy(0)
Properties
• σρx(h) =
1
√
n
if xt is white noise
• σρxy(h) =
1
√
n
if xt or yt is white noise
21.3 Non-Stationary Time Series
Classical decomposition model
xt = µt + st + wt
• µt = trend
• st = seasonal component
• wt = random noise term
21.3.1 Detrending
Least squares
1. Choose trend model, e.g., µt = β0 + β1t + β2t2
2. Minimize rss to obtain trend estimate µt = β0 + β1t + β2t2
3. Residuals noise wt
Moving average
• The low-pass ﬁlter vt is a symmetric moving average mt with aj = 1
2k+1 :
vt =
1
2k + 1
k
i=−k
xt−1
• If 1
2k+1
k
i=−k wt−j ≈ 0, a linear trend function µt = β0 + β1t passes
without distortion
Diﬀerencing
• µt = β0 + β1t =⇒ xt = β1
21.4 ARIMA models
Autoregressive polynomial
φ(z) = 1 − φ1z − · · · − φpzp z ∈ C ∧ φp = 0
Autoregressive operator
φ(B) = 1 − φ1B − · · · − φpBp
Autoregressive model order p, AR (p)
xt = φ1xt−1 + · · · + φpxt−p + wt ⇐⇒ φ(B)xt = wt
AR (1)
• xt = φk
(xt−k) +
k−1
j=0
φj
(wt−j)
k→∞,|φ|<1
=
∞
j=0
φj
(wt−j)
linear process
• E [xt] =
∞
j=0 φj
(E [wt−j]) = 0
• γ(h) = Cov [xt+h, xt] =
σ2
wφh
1−φ2
• ρ(h) = γ(h)
γ(0) = φh
• ρ(h) = φρ(h − 1) h = 1, 2, . . .
24

Moving average polynomial
θ(z) = 1 + θ1z + · · · + θqzq z ∈ C ∧ θq = 0
Moving average operator
θ(B) = 1 + θ1B + · · · + θpBp
MA (q) (moving average model order q)
xt = wt + θ1wt−1 + · · · + θqwt−q ⇐⇒ xt = θ(B)wt
E [xt] =
q
j=0
θjE [wt−j] = 0
γ(h) = Cov [xt+h, xt] =
σ2
w
q−h
j=0 θjθj+h 0 ≤ h ≤ q
0 h > q
MA (1)
xt = wt + θwt−1
γ(h) =



(1 + θ2
)σ2
w h = 0
θσ2
w h = 1
0 h > 1
ρ(h) =
θ
(1+θ2) h = 1
0 h > 1
ARMA (p, q)
xt = φ1xt−1 + · · · + φpxt−p + wt + θ1wt−1 + · · · + θqwt−q
φ(B)xt = θ(B)wt
Partial autocorrelation function (PACF)
• xh−1
i regression of xi on {xh−1, xh−2, . . . , x1}
• φhh = corr(xh − xh−1
h , x0 − xh−1
0 ) h ≥ 2
• E.g., φ11 = corr(x1, x0) = ρ(1)
ARIMA (p, d, q)
d
xt = (1 − B)d
xt is ARMA (p, q)
φ(B)(1 − B)d
xt = θ(B)wt
Exponentially Weighted Moving Average (EWMA)
xt = xt−1 + wt − λwt−1
xt =
∞
j=1
(1 − λ)λj−1
xt−j + wt when |λ| < 1
˜xn+1 = (1 − λ)xn + λ˜xn
Seasonal ARIMA
• Denoted by ARIMA (p, d, q) × (P, D, Q)s
• ΦP (Bs
)φ(B) D
s
d
xt = δ + ΘQ(Bs
)θ(B)wt
21.4.1 Causality and Invertibility
ARMA (p, q) is causal (future-independent) ⇐⇒ ∃{ψj} :
∞
j=0 ψj < ∞ such that
xt =
∞
j=0
wt−j = ψ(B)wt
ARMA (p, q) is invertible ⇐⇒ ∃{πj} :
∞
j=0 πj < ∞ such that
π(B)xt =
∞
j=0
Xt−j = wt
Properties
• ARMA (p, q) causal ⇐⇒ roots of φ(z) lie outside the unit circle
ψ(z) =
∞
j=0
ψjzj
=
θ(z)
φ(z)
|z| ≤ 1
• ARMA (p, q) invertible ⇐⇒ roots of θ(z) lie outside the unit circle
π(z) =
∞
j=0
πjzj
=
φ(z)
θ(z)
|z| ≤ 1
Behavior of the ACF and PACF for causal and invertible ARMA models
AR (p) MA (q) ARMA (p, q)
ACF tails off cuts off after lag q tails off
PACF cuts off after lag p tails off q tails off
21.5 Spectral Analysis
Periodic process
xt = A cos(2πωt + φ)
= U1 cos(2πωt) + U2 sin(2πωt)
• Frequency index ω (cycles per unit time), period 1/ω
25

• Amplitude A
• Phase φ
• U1 = A cos φ and U2 = A sin φ often normally distributed rv’s
Periodic mixture
xt =
q
k=1
(Uk1 cos(2πωkt) + Uk2 sin(2πωkt))
• Uk1, Uk2, for k = 1, . . . , q, are independent zero-mean rv’s with variances σ2
k
• γ(h) =
q
k=1 σ2
k cos(2πωkh)
• γ(0) = E x2
t =
q
k=1 σ2
k
Spectral representation of a periodic process
γ(h) = σ2
cos(2πω0h)
=
σ2
2
e−2πiω0h
+
σ2
2
e2πiω0h
=
1/2
−1/2
e2πiωh
dF(ω)
Spectral distribution function
F(ω) =



0 ω < −ω0
σ2
/2 −ω ≤ ω < ω0
σ2
ω ≥ ω0
• F(−∞) = F(−1/2) = 0
• F(∞) = F(1/2) = γ(0)
Spectral density
f(ω) =
∞
h=−∞
γ(h)e−2πiωh
−
1
2
≤ ω ≤
1
2
• Needs
∞
h=−∞ |γ(h)| < ∞ =⇒ γ(h) =
1/2
−1/2
e2πiωh
f(ω) dω h = 0, ±1, . . .
• f(ω) ≥ 0
• f(ω) = f(−ω)
• f(ω) = f(1 − ω)
• γ(0) = V [xt] =
1/2
−1/2
f(ω) dω
• White noise: fw(ω) = σ2
w
• ARMA (p, q) , φ(B)xt = θ(B)wt:
fx(ω) = σ2
w
|θ(e−2πiω
)|2
|φ(e−2πiω)|2
where φ(z) = 1 −
p
k=1 φkzk
and θ(z) = 1 +
q
k=1 θkzk
Discrete Fourier Transform (DFT)
d(ωj) = n−1/2
n
i=1
xte−2πiωj t
Fourier/Fundamental frequencies
ωj = j/n
Inverse DFT
xt = n−1/2
n−1
j=0
d(ωj)e2πiωj t
Periodogram
I(j/n) = |d(j/n)|2
Scaled Periodogram
P(j/n) =
4
n
I(j/n)
=
2
n
n
t=1
xt cos(2πtj/n
2
+
2
n
n
t=1
xt sin(2πtj/n
2
22 Math
22.1 Gamma Function
• Ordinary: Γ(s) =
∞
0
ts−1
e−t
dt
• Upper incomplete: Γ(s, x) =
∞
x
ts−1
e−t
dt
• Lower incomplete: γ(s, x) =
x
0
ts−1
e−t
dt
• Γ(α + 1) = αΓ(α) α > 1
• Γ(n) = (n − 1)! n ∈ N
• Γ(1/2) =
√
π
22.2 Beta Function
• Ordinary: B(x, y) = B(y, x) =
1
0
tx−1
(1 − t)y−1
dt =
Γ(x)Γ(y)
Γ(x + y)
• Incomplete: B(x; a, b) =
x
0
ta−1
(1 − t)b−1
dt
• Regularized incomplete:
Ix(a, b) =
B(x; a, b)
B(a, b)
a,b∈N
=
a+b−1
j=a
(a + b − 1)!
j!(a + b − 1 − j)!
xj
(1 − x)a+b−1−j
26

• I0(a, b) = 0 I1(a, b) = 1
• Ix(a, b) = 1 − I1−x(b, a)
22.3 Series
Finite
•
n
k=1
k =
n(n + 1)
2
•
n
k=1
(2k − 1) = n2
•
n
k=1
k2
=
n(n + 1)(2n + 1)
6
•
n
k=1
k3
=
n(n + 1)
2
2
•
n
k=0
ck
=
cn+1
− 1
c − 1
c = 1
Binomial
•
n
k=0
n
k
= 2n
•
n
k=0
r + k
k
=
r + n + 1
n
•
n
k=0
k
m
=
n + 1
m + 1
• Vandermonde’s Identity:
r
k=0
m
k
n
r − k
=
m + n
r
• Binomial Theorem:
n
k=0
n
k
an−k
bk
= (a + b)n
Inﬁnite
•
∞
k=0
pk
=
1
1 − p
,
∞
k=1
pk
=
p
1 − p
|p| < 1
•
∞
k=0
kpk−1
=
d
dp
∞
k=0
pk
=
d
dp
1
1 − p
=
1
(1 − p)2
|p| < 1
•
∞
k=0
r + k − 1
k
xk
= (1 − x)−r
r ∈ N+
•
∞
k=0
α
k
pk
= (1 + p)α
|p| < 1 , α ∈ C
22.4 Combinatorics
Sampling
k out of n w/o replacement w/ replacement
ordered nk
=
k−1
i=0
(n − i) =
n!
(n − k)!
nk
unordered
n
k
=
nk
k!
=
n!
k!(n − k)!
n − 1 + r
r
=
n − 1 + r
n − 1
Stirling numbers, 2nd
kind
n
k
= k
n − 1
k
+
n − 1
k − 1
1 ≤ k ≤ n
n
0
=
1 n = 0
0 else
Partitions
Pn+k,k =
n
i=1
Pn,i k > n : Pn,k = 0 n ≥ 1 : Pn,0 = 0, P0,0 = 1
Balls and Urns f : B → U D = distinguishable, ¬D = indistinguishable.
|B| = n, |U| = m f arbitrary f injective f surjective f bijective
B : D, U : ¬D mn mn
m ≥ n
0 else
m!
n
m
n! m = n
0 else
B : ¬D, U : D
n + n − 1
n
m
n
n − 1
m − 1
1 m = n
0 else
B : D, U : ¬D
m
k=1
n
k
1 m ≥ n
0 else
n
m
1 m = n
0 else
B : ¬D, U : ¬D
m
k=1
Pn,k
1 m ≥ n
0 else
Pn,m
1 m = n
0 else
References
[1] P. G. Hoel, S. C. Port, and C. J. Stone. Introduction to Probability Theory.
Brooks Cole, 1972.
[2] L. M. Leemis and J. T. McQueston. Univariate Distribution Relationships.
The American Statistician, 62(1):45–53, 2008.
[3] R. H. Shumway and D. S. Stoﬀer. Time Series Analysis and Its Applications
With R Examples. Springer, 2006.
[4] A. Steger. Diskrete Strukturen – Band 1: Kombinatorik, Graphentheorie,
Algebra. Springer, 2001.
[5] A. Steger. Diskrete Strukturen – Band 2: Wahrscheinlichkeitstheorie und
Statistik. Springer, 2002.
[6] L. Wasserman. All of Statistics: A Concise Course in Statistical Inference.
Springer, 2003.
27

Univariatedistributionrelationships,courtesyLeemisandMcQueston[2].
28

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