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Scripts were run on a Xeon E3-1245v3 @ 3.40GHz (1/4 cores) 16GB RAM with 4.2.1 on Windows 7 build 7601, Service Pack 1, Universal C Runtime 10.0.10240.16390.

- The right-hand badges give the respective section’s ‘level’.

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- An advanced knowledge of PK and/or R is required. Not recommended for beginners in particular.

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**ne**R**d**or PK afficionado, skipping is recommended. Suggested for experts but might be confusing for others.

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Which method should we use to demonstrate achievement of steady state?

First of all, design the study based on a *worst-case* half
life (__not__ an average).

Especially for prolonged (aka
controlled, sustained, extended, long-acting) release formulations
consider flip-flop Pharmacokinetics.

“You can’t fix by analysis what you bungled by design.

For a comparative bioavailability study we have essentially two design options:

- Single dose
- Multiple dose

The former is recommended in all guidelines for comparing immediate
release formulations because it is more sensitive in detecting potential
differences in formulations (especially of the rate of absorption \(\small{C_\textrm{max}}\) / \(\small{t_\textrm{max}}\)). Only in rare
cases (*e.g.*, an auto-inducer like carbamazepine^{4}) studies after
multiple doses are more sensitive than after a single dose.

The latter is required in studies in patients, where the treatment
cannot be interrupted with washout phases (*e.g.*, in oncology).
It is mandatory for the
EMA when comparing
prolonged release formulations – *additionaly* to the single dose
study.

“In steady-state studies, wash-out of the last dose of the previous treatment can overlap with the build-up of the second treatment, provided the build-up period is sufficiently long (at least three times the dominating half-life).

“Whenever multiple dose studies are performed it should be demonstrated that steady state has been reached.

“In steady-state studies wash out of the previous treatment last dose can overlap with the build-up of the second treatment, provided the build-up period is sufficiently long (at least three times the terminal half-life).

A similar wording is used by the
WHO,^{8} only that »at least
five times the terminal half-life« are recommended.

In Europe regularly a linear regression of the last three pre-dose concentrations was performed, allowing to exclude subjects who are not in steady state. The idea behind is based on elementary pharmacokinetics. By definition in steady state the rate of inflow equals the rate of outflow. Hence, in all dosing intervals the pre-dose concentrations (\(\small{C_{\textrm{pd}}}\)) are expected to be similar. Therefore, a linear regression should give a slope close to zero. However, this approach was based on custom and not backed by a guideline.

Recall the accumulation index \(\small{R=1/(1-2^{-\tau\,/\,t_{1/2}}\,)}\).
\(\small{R}\) is strictly valid for
intravenous doses and approximate for extravascular administrations in a
one-compartment model if \(\small{k_{01}\gg
k_{10}}\).

An example with \(\small{\tau=t_{1/2}\;(R=2)}\), where \(\small{C_\textrm{pd}}\) is the pre-dose
concentration and \(\small{C_\tau}\) is
the concentration at the end of the dosing interval. Values ‘by
definition’ are denoted by \(\small{\ast}\). \[\small{
\begin{array}{rrrrrr}
\text{dose} & \text{time} & C_\textrm{pd}\phantom{00000} &
\text{% change} & \text{moving slope} & C_\tau\phantom{000000}\\
\hline
1 & 0 & 0\phantom{.000000}\phantom{0} & - & - &
50\phantom{.00000000}\\
2 & 24 & 50\phantom{.0000000} & 50\phantom{.0000000} & -
& 75\phantom{.00000000}\\
3 & 48 & 75\phantom{.0000000} & 16.6666667 & 0.78125
& 87.5\phantom{0000000}\\
4 & 72 & 87.5\phantom{000000} & 7.1428571 & 0.39063
& 93.75\phantom{000000}\\
5 & 96 & 93.75\phantom{00000} & 3.3333333 & 0.19531
& 96.875\phantom{00000}\\
6 & 120 & 96.875\phantom{0000} & 1.6129032 & 0.09766
& 98.4375\phantom{0000}\\
7 & 144 & 98.4375\phantom{000} & 0.7936508 & 0.04883
& 99.21875\phantom{000}\\
8 & 168 & 99.21875\phantom{00} & 0.3937008 & 0.02441
& 99.609375\phantom{00}\\
9 & 192 & 99.609375\phantom{0} & 0.1960784 & 0.01221
& 99.8046875\phantom{0}\\
10 & 216 & 99.8046875 & 0.0978474 & 0.00610 &
99.90234375\\
\infty & \infty & 100\phantom{.0000000} &
0\ast\phantom{.00000} & 0\ast\phantom{.000} &
100\;(=C_\textrm{pd})\phantom{0}\\
\hline
\end{array}}\]

Health Canada required 1992 – 2011 multiple dose studies for drugs
which accumulated, *i.e.*, if the extrapolated \(\small{AUC}\) after a single dose was
> 20% of \(\small{AUC_{0-\infty}}\).

“7.6.2 Evidence of Steady State

For steady-state studies of drugs with uncomplicated characteristics, at least three consecutive pre-dose concentration levels (C_{pd}) are required to provide evidence of steady state. Generally, observations of C_{pd}for the test and reference products should be recorded at the same time of the day. One of these measurements could be taken based on the first sample of the study day in which a profile over the dosing interval is being established. Steady state is usually achieved when repeated doses of a formulation are administered over a period that exceeds five disposition half-lives of the modified-release form.

*‘Evidence of steady state’* had to be assessed by a repeated
measures analysis with an amazing number of 11 (__eleven‼__) factors.
If you want to give it a try, download the
data compiled from the guidance’s Table 12-A (Randomization), Table 12-G (Test), and Table 12-H (Reference).

```
<- read.csv("https://bebac.at/downloads/HC1996-TABLE-12.csv",
data colClasses = c(rep("factor", 5), "integer"))
<- function(x) {
geo.mean <- as.numeric(unlist(unique(x[1])))
t <- as.integer(unlist(unique(x[2])))
n <- NA_real_
gm for (i in seq_along(t)) {
<- x[x[1] == t[i], ]
y if (sum(y$C <= 0) > 0) {
<- y[-which(y$C <= 0), ]
y
}if (nrow(y) > 0) {
<- exp(mean(log(y$C), na.rm = TRUE))
gm[i]
}
}return(gm)
}<- unique(as.character(data$Subject))
Subject <- unique(as.character(data$Formulation))
Form <- data.frame(Formulation = rep(Form, each = length(Subject)),
slopes Subject = Subject, slope = NA_real_,
lower.CL = NA_real_, upper.CL = NA_real_,
sig = FALSE)
<- data$Time
t <- as.numeric(levels(t))[t]
t <- range(data$C.pd)
ylim <- range(t)
xlim <- c("blue", "red")
cols <- colorRampPalette(cols)(length(Subject))
col <- paste0(col, "80")
col <- par(no.readonly=TRUE)
op par(mar = c(4.1, 4, 0.2, 0.2), cex.axis = 0.9)
split.screen(c(2, 1))
for (i in seq_along(Form)) {
screen(i)
ifelse (i == 1, xlab <- "", xlab <- "Time (day)")
plot(xlim, ylim, type = "n", las = 1, xlab = xlab,
ylab = paste0("C.pd (", Form[i], ")"))
grid()
<- data.frame(t = t,
x C = data[data$Formulation == Form[i], 6])
lines(unique(t), geo.mean(x), lwd = 2)
for (j in seq_along(Subject)) {
<- data$Time[data$Formulation == Form[i] &
t $Subject == Subject[j]]
data<- as.numeric(levels(t))[t]
t <- data$C.pd[data$Formulation == Form[i] &
C $Subject == Subject[j]]
datapoints(t, C, pch = 16, cex = 1.1, col = col[j])
<- data.frame(t = t, C = C)
x <- tail(x, 3)
x <- lm(C ~ t, data = x)
m lines(x$t, predict(m), col = substr(col[j], 1, 7))
$slope[slopes$Formulation == Form[i] &
slopes$Subject == Subject[j]] <- coef(m)[2]
slopes<- suppressWarnings(confint(m, "t"))
ci $Formulation == Form[i] &
slopes[slopes$Subject == Subject[j], 4:5] <- sprintf("%+.4f", ci)
slopesif (sign(ci[1]) == sign(ci[2]))
$sig[slopes$Formulation == Form[i] &
slopes$Subject == Subject[j]] <- TRUE
slopes
}
}close.screen(all = TRUE)
par(op)
$slope <- sprintf("%+5.1f", slopes$slope)
slopesnames(slopes)[6] <- "sign # 0"
.3 <- data[as.numeric(as.character(data$Time)) >= 6, ]
last<- lm(log(C.pd) ~ Sequence + Subject * Sequence + Period +
muddle + Subject * Formulation * Sequence +
Formulation + Time * Sequence + Time * Subject * Sequence +
Time * Period + Time * Period +
Time * Formulation * Subject * Sequence,
Time data = last.3)
print(slopes, row.names = FALSE); anova(muddle) # Oops!
```

“Of main importance in this analysis is the time and time*form interaction. The error term to test time is the time*ID(Seq). The error term to test time*form is time*form*ID(Seq). Should either the time or time*form effects be significant, the study may not have been at steady state for one or both formulations.

Back in the day I failed to reproduce the results given in Table 12-I.

In `R`

I got …

```
Analysis of Variance Table
Response: log(C.pd)
Df Sum Sq Mean Sq F value Pr(>F)
Sequence 1 0.01326 0.013260 NaN NaN
Subject 10 1.22059 0.122059 NaN NaN
Period 1 0.01825 0.018249 NaN NaN
Formulation 1 0.22069 0.220686 NaN NaN
Time 2 0.11935 0.059675 NaN NaN
Subject:Formulation 10 0.56173 0.056173 NaN NaN
Sequence:Time 2 0.13041 0.065204 NaN NaN
Subject:Time 20 0.71386 0.035693 NaN NaN
Period:Time 2 0.06441 0.032204 NaN NaN
Formulation:Time 2 0.05961 0.029804 NaN NaN
Subject:Formulation:Time 20 0.60824 0.030412 NaN NaN
Residuals 0 0.00000 NaN
Warning message:
In anova.lm(m) : ANOVA F-tests on an essentially perfect fit are unreliable
```

… and in Phoenix WinNonlin (all effects fixed):

`ERROR 11068: There are no degrees of freedom for residual. No tests are possible.`

When I specified all subject-related effects as random:

```
Hypothesis Numer_DF Denom_DF F_stat P_value
-----------------------------------------------------------------
int 1 60 4446.56 0.0000
Sequence 1 60 0.0609162 0.8059
Formulation 1 60 3.04158 0.0863
Time 2 60 1.00778 0.3711
Time*Sequence 2 60 1.19822 0.3088
Time*Period 3 60 1.59434 0.2002
Time*Formulation 2 60 1.64308 0.2020
```

\[\small{\text{TABLE 12-I }\textsf{(in all
of its doubtful beauty)}}\] \[\small{\begin{array}{lrcccc}
\hline\text{Source} & \text{df} & \text{SS} & \text{MS}
& \text{F} & \text{PR > F}\\\hline
\text{Seq} & 1 & 0.01500416 & 0.01500416 & 0.12 &
0.7313\\
\text{ID(Seq)} & 10 & 1.20256303 & 1.20256303 & 2.11
& 0.1278\\
\text{Period} & 1 & 0.04556284 & 0.04556284 & 0.80 &
0.3926\\
\text{Form} & 1 & 0.22620312 & 0.22620312 & 3.96 &
0.0745\\
\text{ID*Form(Seq)} & 10 & 0.57066567 & 0.0570666\phantom{0}
& - & -\\
\text{Time} & 2 & 0.11182757 & 0.05591378 & 1.60 &
0.2274\\
\text{Time*Seq} & 10 & 0.13927056 & 0.06963528 & 1.99
& 0.1631\\
\text{Time*ID(Seq)} & 20 & 0.70058566 & 0.03502928 & -
& -\\
\text{Time*Period} & 2 & 0.07672565 & 0.03836282 & 1.24
& 0.3110\\
\text{Time*Form} & 2 & 0.06390201 & 0.03195101 & 1.03
& 0.3746\\
\text{Time*Form*ID(Seq)} & 20 & 0.61937447 & 0.03096872
& - & -\\
\text{Corrected Total} & 71 & 3.74093818 & - & - &
-\\\hline
\end{array}}\] Maybe you are equipped with Harry Potter’s magic
wand. If you succeed in reproducing this table, drop me a note
*how* you did it. I’m always eager to learn.

However, the outcome of such an analysis is dichotomous (pass |
fail), risking to discard the entire [*sic*] study. The family-wise
error rate of two simultaneous tests at level \(\small{\alpha}\) is given by \[FWER=1-(1-\alpha)^2
\tag{1}\] or 9.75% for \(\small{\alpha=0.05}\). That means,
following this method you may have to trash almost one of ten studies
*although* steady state was achieved.

Given, that is only of historical interest.

“Achievement of steady state can be evaluated by collecting pre-dose samples on the day before the PK assessment day and on the PK assessment day. A specific statistical method to assure that steady state has been reached is not considered necessary in bioequivalence studies. Descriptive data is sufficient.

The requirement of performing multiple dose studies was lifted in
2012 by Health Canada. If such studies are performed, the pre-dose
concentrations have to be reported.^{11}

“Because single-dose studies are considered more sensitive in addressing the primary question of BE (e.g., release of the drug substance from the drug product into the systemic circulation), multiple-dose studies are generally not recommended.

A statistical method is not given in any guidance of the
FDA. A statement
about the current *practice*:^{13}

“Current FDA practice for general procedure for steady-state assessment is by comparing at least three pre-dose concentrations for each formulation using linear regression techniques to evaluate achievement of steady state and by comparing the BE outcome with or without the subjects whose terminal [sic] slope is significantly different from 0 (i.e., they did not attain steady state).

“Whenever multiple dose studies are performed it should be demonstrated that steady state has been reached. Achievement of steady-state is assessed by comparing at least three pre-dose concentrations for each formulation, unless otherwise justified.

Since the EMA’s
guideline is adopted in Australia, the same story Down Under.^{15}

While appealing,^{16} ^{17} a regression approach is problematic.

- In any regression it is assumed that responses are
*not*correlated (*i.e.*, depend only on the predictor). In PK clearly this is not the case – pre-dose concentrations are*highly*correlated. Perform a Durbin–Watson test – you will be surprised. - The outcome depends on the within-subject variability:
- If it is low to intermediate, it is quite likely that some subjects
show a slope which is statistically significant different from zero (or
alternatively, its confidence interval (CI) does not contain
zero).

It will be – falsely – concluded that the respective subjects are not in steady-state. - If it is high, a significant difference of the slope to zero
(
*i.e.*, the CI contains zero) will not be detected – although some subjects are still in the build-up phase.

It will be – falsely – concluded that steady-state is achieved.

- If it is low to intermediate, it is quite likely that some subjects
show a slope which is statistically significant different from zero (or
alternatively, its confidence interval (CI) does not contain
zero).

For every subject a linear model is set up \[Y=a+bX+\epsilon\small{\textsf{,}}\tag{2}\]
where \(\small{Y}\) is the vector of
responses (here the pre-dose concentrations), \(\small{X}\) the vector of predictors (here
the pre-dose sampling time points), and \(\epsilon\) the unexplained error with \(\small{\mathcal{N}(0;\sigma^2)}\). Fitting
the data gives estimates of the intercept \(\small{\hat{a}}\) and slope \(\small{\hat{b}}\). One can either perform a
one-sample
*t*-test of \(\small{\hat{b}}\) versus zero with \(\small{\alpha=0.05}\) (significantly
different if \(\small{p(t)<0.05}\))
or assess whether zero is contained in the 95%
CI. Since the latter is more
‘popular’, we employed it in the following.

The script is for a one-compartment model, first order absorption and elimination with an optional lag-time. Flip-flop PK (\(\small{k_{01}=k_{10}}\)) is implemented. Cave: 288 LOC.

```
<- function(nsims, f, D, V, t12.a, t12.e, tlag, tau, fact,
sim.ss CV.lo = 0.1, CV.hi = 0.4, delta = 0.2,
method = "linear", dw = FALSE, setseed = TRUE,
progress = FALSE, plots = FALSE) {
suppressMessages(library(car)) # for Durbin-Watson
# Cave: Long run time if nsims > 1e4
# Extremely long run time if dw = TRUE
if (nsims < 1) stop("nsims must be at least 1.")
if (f <= 0 | f > 1) stop("f must be positive and <= 1.")
if (D <= 0) stop("D must be positive.")
if (t12.a <= 0) stop("t12.a (absorption half life) must be positive.")
if (t12.e <= 0) stop("t12.e (elimination half life) must be positive.")
if (tlag < 0) stop("tlag must >= 0.")
if (tau < 0) stop("tau must be positive.")
if (CV.lo <= 0) stop("CV.lo must be positive.")
if (CV.hi <= 0) stop("CV.hi must be positive.")
if (delta < 0) stop("delta must be positive.")
if (fact < 3) {
message("fact < 3 not reasonable. Forced to 3.")
<- 3
fact
}if (!method %in% c("linear", "power")) stop("method not supported.")
<- function(f, D, V, k01, k10, tlag, t) {
one.comp # one-compartment model, first order absorption
# and elimination
if (!isTRUE(all.equal(k01, k10))) {# common: k01 != k10
<- f * D * k01 / (V * (k01 - k10)) *
C exp(-k10 * (t - tlag)) - exp(-k01 * (t - tlag)))
(<- log(k01 / k10) / (k01 - k10) + tlag
tmax <- f * D * k01 / (V * (k01 - k10)) *
Cmax exp(-k10 * tmax) - exp(-k01 * tmax))
(else { # flip-flop
}<- k10
k <- f * D / V * k * (t - tlag) * exp(-k * (t - tlag))
C <- 1 / k
tmax <- f * D / V * k * tmax * exp(-k * tmax)
Cmax
}<= 0] <- 0 # correct negatives caused by lag-time
C[C <- list(C = C, Cmax = Cmax, tmax = tmax)
res return(res)
}
<- function(x, y) {
round.up # round x up to the next multiple of y
return(as.integer(y * (x %/% y + as.logical(x %% y))))
}
<- function(f, D, V, k01, k10, tlag, tau, t, n.D) {
superposition # single dose
<- data.frame(adm = 1L, dose = NA, t = t,
x C = one.comp(f, D, V, k01, k10, tlag, t)$C)
$dose[1] <- D
x# multiple dose (nonparametric superposition)
<- x
y for (i in 1:nrow(y)) {
if (isTRUE(all.equal(y$t[i] %% tau, 0))) y$dose[i] <- D
}<- unique(diff(y$t))
delta <- seq(0, max(y$t), tau) # administration times
adm <- 1:length(adm) # number of administration
adm.n if (isTRUE(all.equal(delta, tau))) {# for pre-dose concentrations
$adm <- adm.n
yelse { # for build-up
}<- sum(y$t < tau)
t.tau $adm <- head(rep(adm.n, each = t.tau), nrow(y))
y
}<- 2:(n.D + 1)
z for (i in seq_along(z)) {
<- y$C[y$adm >= z[i]]
tmp.y <- x$C[1:length(tmp.y)]
tmp.x $C[y$adm >= z[i]] <- tmp.y + tmp.x
y
}<- list(SD = x, MD = y)
res return(res)
}
<- function(CV) {
CV2var # convert CV to variance
return(log(CV^2 + 1))
}
<- function(x, method = "linear", dw = FALSE, delta = 0.2) {
mods # last three pre-dose concentrations
<- tail(x, 3)
x # model and 95% CI of slope
if (method == "linear") {
<- lm(C ~ t, data = x)
m <- coef(m)[[1]]
a <- confint(m, "t", level = 0.95)
ci if (dw) dw.p <- dwt(m, reps = 1e4)$p
}if (method == "power") {
<- lm(log(C) ~ log(t), data = x)
m <- exp(coef(m)[[1]])
a <- confint(m, "log(t)", level = 0.95)
ci if (dw) dw.p <- dwt(m, reps = 1e4)$p
}# assess whether the CI contains zero
ifelse (sign(ci[1]) == sign(ci[2]), sig1 <- TRUE, sig1 <- FALSE)
# assess whether the CI is within +/- delta
ifelse ((ci[1] >= -delta & ci[2] <= +delta), sig2 <- FALSE, sig2 <- TRUE)
<- list(method = method, a = a, b = coef(m)[[2]],
res lwr = ci[[1]], upr = ci[[2]], sig1 = sig1, sig2 = sig2,
t.start = x$t[1], t.end = x$t[3], dw.p = dw.p)
return(res)
}
<- function(x) {
geo.mean # geometric mean
<- as.integer(unlist(unique(x[1])))
n <- as.numeric(unlist(unique(x[2])))
t <- NA_real_
gm for (i in seq_along(t)) {
<- x[x[2] == t[i], ]
y if (sum(y$C <= 0) > 0) {
<- y[-which(y$C <= 0), ] # only positive values
y
}if (nrow(y) > 0) {
<- exp(mean(log(y$C), na.rm = TRUE))
gm[i]
}
}return(gm)
}
if (nsims > 1e4) progress <- TRUE # to keep the spirit high
<- log(2) / t12.a # absorption rate constant
k01 <- log(2) / t12.e # elimination rate constant
k10 ifelse (t12.e >= t12.a, t.last <- fact * t12.e, t.last <- fact * t12.a)
<- round.up(t.last, tau) # time of last dose
t.last <- seq(0, t.last, tau) # time of administrations
t <- length(t) # number of doses
n.D if (setseed) set.seed(1234567) # for reproducibility
# build-up (high resolution)
<- seq(0, t.last, length.out = (t.last * 10 + 1))
t.bu <- superposition(f, D, V, k01, k10, tlag, tau, t = t.bu, n.D)$MD
bu # pre-dose concentrations
<- bu[!is.na(bu$dose), ]
MD <- mods(MD, method, dw)$a # intercept without error
mod.a <- mods(MD, method, dw)$b # slope/exponent without error
mod.b <- tail(MD$C, 1) # last pre-dose of model
C.ss # simulated pre-dose concentrations
<- sim.hi <- data.frame(sim = rep(1:nsims, each = n.D),
sim.lo t = t, C = NA_real_)
# results with low and high variability
<- hi <- data.frame(sim = 1:nsims, method = NA_character_,
lo a = NA_real_, b = NA_real_,
lwr = NA_real_, upr = NA_real_,
sig1 = NA, sig2 = NA,
t.start = NA_real_, t.end = NA_real_)
if (progress) pb <- txtProgressBar(0, 1, 0, char = "\u2588",
width = NA, style = 3)
for (sim in 1:nsims) {
# multiplicative error
<- rlnorm(n = n.D,
C.lo meanlog = log(MD$C) - 0.5 * CV2var(CV.lo),
sdlog = sqrt(CV2var(CV.lo)))
<- rlnorm(n = n.D,
C.hi meanlog = log(MD$C) - 0.5 * CV2var(CV.hi),
sdlog = sqrt(CV2var(CV.hi)))
$C[sim.lo$sim == sim] <- C.lo
sim.lo$C[sim.hi$sim == sim] <- C.hi
sim.hi# get slope, 95% CI, and significance
2:11] <- mods(sim.lo[sim.lo$sim == sim, ], method, dw)
lo[sim, 2:11] <- mods(sim.hi[sim.lo$sim == sim, ], method, dw)
hi[sim, if (progress) setTxtProgressBar(pb, sim / nsims)
# simulations running...
}if (progress) close(pb)
if (method == "linear") {
<- "slope"
txt <- "linear model: C ~ a + b * t"
me else {
}<- "exponent"
txt <- "power model: C ~ a * t^b"
me
}<- paste(me, "\n delta =", delta,
me "(acceptance limits {-delta, +delta})\n")
# cosmetics
<- paste0("CV = %4.1f%%, median of ", txt, "s = %+.5f")
fm <- paste0(fm, "\n %5.1f%% positive ", txt, "s")
fm <- paste0(fm, "\n %5.1f%% negative ", txt, "s")
fm <- paste0(fm, "\n significant ", txt, "s: %6.2f%% %s")
fm <- paste0(fm, "\n failed ", txt, "s : %6.2f%% %s")
fm # results
<- paste(prettyNum(nsims, format = "d", big.mark =","), "simulations,",
txt "of model =", sprintf("%+.5f", mod.b),
me, txt, "\n last model’s pre-dose concentration =", signif(C.ss, 4), "\n",
sprintf(fm, 100 * CV.lo, median(lo$b),
100 * sum(lo$b > 0) / nsims,
100 * sum(lo$b < 0) / nsims,
100 * sum(lo$sig1, na.rm = TRUE) / nsims,
"(95% CI does not contain zero)",
100 * sum(lo$sig2, na.rm = TRUE) / nsims,
"(95% CI not within acceptance limits)\n"))
if (dw) {
<- paste(txt, " autocorrelation :",
txt sprintf("%5.1f%%",
100 * sum(lo$dw.p < 0.05, na.rm = TRUE) / nsims), "\n")
}<- paste(txt, sprintf(fm, 100 * CV.hi, median(hi$b),
txt 100 * sum(hi$b > 0) / nsims,
100 * sum(hi$b < 0) / nsims,
100 * sum(hi$sig1, na.rm = TRUE) / nsims,
"(95% CI does not contain zero)",
100 * sum(hi$sig2, na.rm = TRUE) / nsims,
"(95% CI not within acceptance limits)\n"))
if (dw) {
<- paste(txt, " autocorrelation :",
txt sprintf("%5.1f%%",
100 * sum(hi$dw.p < 0.05, na.rm = TRUE) / nsims), "\n")
}cat(txt)
if (plots) {
# use [Page][up] / [Page][down] keys to switch
# between plots in the graphics device
windows(width = 4.5, height = 4.5, record = TRUE)
<- par(no.readonly = TRUE)
op par(mar = c(4, 3.9, 0.2, 0.1), cex.axis = 0.9)
<- range(bu$C)
ylim if (method == "power") {
.3 <- tail(MD$t, 3)
last<- seq(last.3[1], last.3[3], length.out = 251)
t.tmp
}for (sim in 1:nsims) {
if (sim == 1) {
plot(bu$t, bu$C, type = "l", ylim = ylim, axes = FALSE,
xlab = "time (h)", ylab = "concentration (unit)")
grid(nx = NA, ny = NULL)
abline(v = t, lty = 3, col = "lightgrey")
axis(1, at = t)
axis(2, las = 1)
box()
points(MD$t, MD$C, pch = 15, cex = 1.25)
}points(sim.lo$t[sim.lo$sim == sim], sim.lo$C[sim.lo$sim == sim],
pch = 21, cex = 1.1, col = "blue", bg = "#A0A0FF")
if (lo$sig1[sim]) {
<- "red"
col <- 2
lwd else {
}<- "#A0A0FF"
col <- 1
lwd
}if (method == "linear") {
lines(x = lo[sim, 9:10],
y = c(lo$a[sim] + lo$b[sim] * lo$t.start[sim],
$a[sim] + lo$b[sim] * lo$t.end[sim]), col = col, lwd = lwd)
lo
}if (method == "power") {
lines(x = t.tmp, y = lo$a[sim] * t.tmp^lo$b[sim], col = col, lwd = lwd)
}
}points(t, geo.mean(sim.lo), pch = 24, cex = 1.5, bg = "#FF0808BF")
for (sim in 1:nsims) {
if (sim == 1) {
plot(bu$t, bu$C, type = "l", ylim = ylim, axes = FALSE,
xlab = "time (h)", ylab = "concentration (unit)")
grid(nx = NA, ny = NULL)
abline(v = t, lty = 3, col = "lightgrey")
axis(1, at = t)
axis(2, las = 1)
box()
points(MD$t, MD$C, pch = 15, cex = 1.25)
}points(sim.hi$t[sim.hi$sim == sim], sim.hi$C[sim.hi$sim == sim],
pch = 21, cex = 1.1, col = "blue", bg = "#A0A0FF")
if (hi$sig1[sim]) {
<- "red"
col <- 2
lwd else {
}<- "#A0A0FF"
col <- 1
lwd
}if (method == "linear") {
lines(x = hi[sim, 9:10],
y = c(hi$a[sim] + hi$b[sim] * hi$t.start[sim],
$a[sim] + hi$b[sim] * hi$t.end[sim]), col = col, lwd = lwd)
hi
}if (method == "power") {
lines(x = t.tmp, y = hi$a[sim] * t.tmp^hi$b[sim], col = col, lwd = lwd)
}
}points(t, geo.mean(sim.hi), pch = 24, cex = 1.5, bg = "#FF0808BF")
par(op)
} }
```

Let us perform some simulations.

We explore low (\(\small{CV=10\%}\)) and high within-subject
variability (\(\small{CV=40\%}\)). In a
conservative manner we dose for seven half lives.

Additionally to testing whether the
CI contains zero, in analogy to
BE we test also whether it lies
entirely within – arbitrary – acceptance limits of \(\small{\{-\Delta,+\Delta\}}\).

```
<- 1e4L # number of simulations
nsims <- 200L # dose
D <- 2/3 # fraction absorbed (BA)
f <- 3 # volume of distribution
V <- 0.75 # absorption half life
t12.a <- 12 # elimination half life
t12.e <- 0.25 # lag time
tlag <- 24L # dosing interval
tau <- 7L # half life factor
fact <- 0.10 # drug with low variability
CV.lo <- 0.40 # drug with high variability
CV.hi <- 0.2 # for acceptance limits = {1 - delta, 1 + delta}
delta <- TRUE # Assess autocorrelation
dw sim.ss(nsims, f, D, V, t12.a, t12.e, tlag, tau, fact,
dw = dw) CV.lo, CV.hi, delta,
```

```
# 10,000 simulations, linear model: C ~ a + b * t
# delta = 0.2 (acceptance limits {-delta, +delta})
# slope of model = +0.01957
# last model’s pre-dose concentration = 15.97
# CV = 10.0%, median of slopes = +0.01977
# 66.6% positive slopes
# 33.4% negative slopes
# significant slopes: 5.58% (95% CI does not contain zero)
# failed slopes : 78.27% (95% CI not within acceptance limits)
# autocorrelation : 31.9%
# CV = 40.0%, median of slopes = +0.02072
# 54.8% positive slopes
# 45.2% negative slopes
# significant slopes: 4.18% (95% CI does not contain zero)
# failed slopes : 96.75% (95% CI not within acceptance limits)
# autocorrelation : 24.2%
```

With a half life of 12 hours and once daily dosing after five
administrations we are *close* to the true steady state (≈16).
Hence, the slope of the build-up based on the model is slightly
positive. Only in true steady state the slope would be zero (see above). Even then we would expect 5%
significant results by pure chance. That’s the False Positive
Rate (FPR) of any test performed at level \(\small{\alpha=0.05}\).

With low variability we are above the
FPR. That’s correct because the
*true* slope *is* positive indeed.

On the other hand, with high variability the test is not powerful enough
to detect the positive slope and we are close to the level of the
test.

Note also that with low variability we see more positive slopes than negative ones – which is correct in principle. With high variability everything drowns in noise.

The confidence interval inclusion approach fails with flying colors. Definitely not a good idea.

Let’s explore a smaller subset, which reflects better what we might face in a particular study.

Here with low variability three simulated subjects show a significant slope (FPR 12.5%) and with high variability none. When assessing the Durbin–Watson test, we find a significant autocorrelation in six simulated subjects with low variability and in five with high variability. Hence, it confirms that applying a regression is not justified.

Of course, in reality the level of the true steady state is unknown. The geometric means show that the pre-dose concentrations are stable, though with low variability they seem to rise a bit.

A shorter build-up (fewer doses) might be risky. Here planned with five half lives as recommended in the guidelines (only four doses, last dosing time rounded up from 60 to 72 hours).

```
10,000 simulations, linear model: C ~ a + b * t
delta = 0.2 (acceptance limits {-delta, +delta})
slope of model = +0.07828
last model’s pre-dose concentration = 15.78
CV = 10.0%, median of slopes = +0.07712
97.3% positive slopes
2.7% negative slopes
significant slopes: 9.46% (95% CI does not contain zero)
failed slopes : 85.79% (95% CI not within acceptance limits)
autocorrelation : 6.2%
CV = 40.0%, median of slopes = +0.06995
68.9% positive slopes
31.1% negative slopes
significant slopes: 4.98% (95% CI does not contain zero)
failed slopes : 96.26% (95% CI not within acceptance limits)
autocorrelation : 4.4%
```

The positive slopes are detected. With low variability 9.46% is clearly above the level of the test. With high variability we are close to the level of the test.

As before, we assess a small subset.

With low variability none of the simulated subjects show a significant slope and with high variability one. However, what might a regulatory assessor conclude? With low variability all slopes are positive and with high variability still 19.

An example from my files. A
CNS-drug with polymorphic
metabolism (\(\small{CV_\textrm{inter}\gg
CV_\textrm{intra}}\)). Given the wide range of half lives
reported in the literature and observed in our single dose studies, we
dosed o.a.d. for 14 days, translating
into an average of 15.5 half lives (minimum 5.2, maximum 46). \[\small{\textrm{PK metric (ng/mL):
}\bar{x}_\textrm{geom}\;(CV)}\]\[\small{
\begin{array}{lccc}
\textrm{Treatment} & C_\textrm{pd} & C_\textrm{min} &
C_\tau\\\hline
\textrm{Test} & 28.47\;(80\%) & 26.99\;(79\%) &
33.41\;(59\%)\\
\textrm{Reference} & 29.80\;(77\%) & 28.19\;(75\%) &
32.84\;(61\%)\\\hline
\end{array}}\]\(\small{C_\textrm{pd}>C_\textrm{min}}\),
since in 26 of the 56 profiles we observed a lag time. Recall the
accumulation example above. It is
common that \(\small{C_\tau>C_\textrm{pd}}\) because
we can *never* attain *true* steady state.

Although not assessed in the study, all of these PK metrics would pass conventional bioequivalence. \[\small{\begin{array}{lccccc} \textrm{PK metric} & PE & 90\%\;CL_\textrm{lo} & 90\%\;CL_\textrm{hi} & CV_\textrm{intra} & CV_\textrm{inter}\\\hline \phantom{000}C_\textrm{pd} & \phantom{1}95.54\% & 81.29\% & 112.29\% & 39.5\% & 133.3\%\\ \phantom{000}C_\textrm{min} & \phantom{1}95.74\% & 81.82\% & 112.03\% & 38.3\% & 128.2\%\\ \phantom{000}C_\tau & 100.48\% & 90.27\% & 111.85\% & 25.7\% & \phantom{1}94.3\%\\\hline \end{array}}\] Here the results of the first period.

```
linear model: C ~ a + b * t
Δ = 0.5 (acceptance limits {–Δ, +Δ})
42.9% positive slopes
57.1% negative slopes
significant slopes: 3.57% (95% CI does not contain zero)
failed slopes : 82.14% (95% CI not within acceptance limits)
```

The one subject with a significant slope (with a 95%
CI of 0.0039 – 1.038 it was a
close shave) had low variability. However, at the end of the profile day
the concentration was lower than at 312 hours. Hence, it is reasonable
to conclude that the subject was in steady state and the apparently
rising concentrations were due to pure chance. There was yet another
subject with a similar slope. Since the variability was higher, the
slope was not significant.

Note that ≈82% would fail even with an extremely liberal \(\small{\Delta=0.5}\).

According to the protocol we reported only individual data,
descriptive statistics,^{10} individual and
spaghetti plots. All European regulatory assessors were fine with
that.

Interlude: Flip-flop Pharmacokinetics

\(\small{k_{01}}\): absorption rate
constant

\(\small{k_{10}}\): elimination rate constant

\(\small{k_{10}}\): elimination rate constant

Even in a simple one-compartment model \(\small{k_{01}}\) and \(\small{k_{10}}\) are not identifiable.
Where for IR formulations it is
reasonable to assume \(\small{k_{10}\ll
k_{01}}\), *i.e.*, that absorption is much faster than
elimination, for prolonged release formulations this must not be – and
very likely is not – the case. If you are a dinosaur of biopharmaceutics
like me, perhaps you remember the slogan of the 1980s »*the flatter is better*«. A main target of
product development is to decrease \(\small{k_{01}}\) (increase the absorption
half life). Never plan the study based on the apparent elimination half
life of an IR formulation.

Here an example of flip-flop PK
with \(\small{k_{01}=k_{10}=\textsf{8
h}}\), designed with the *naïve* dosing of seven
elimination half lives (56 rounded up to 72 hours).

As usual: Know the drug and know your formulation!

Given the behavior of the build-up, which is concave, one might be
tempted set up a power model \[Y=a\cdot
X^{\,b}+\epsilon\small{\textsf{.}}\tag{3}\] Like in
dose-proportionality^{18} for convenience its linearized form \[\log_{e}Y=\log_{e}a+b\cdot\log_{e}X+\epsilon
\tag{4}\] can be fitted. The tests are the same like in
the linear model \(\small{(2)}\).

```
<- 1e4L # number of simulations
nsims <- 200L # dose
D <- 2/3 # fraction absorbed (BA)
f <- 3 # volume of distribution
V <- 0.75 # absorption half life
t12.a <- 12 # elimination half life
t12.e <- 0.25 # lag time
tlag <- 24L # dosing interval
tau <- 5L # half life factor
fact <- 0.10 # drug with low variability
CV.lo <- 0.40 # drug with high variability
CV.hi <- 0.2 # for acceptance limits = {1 - delta, 1 + delta}
delta <- TRUE # Assess autocorrelation
dw <- "power"
method sim.ss(nsims, f, D, V, t12.a, t12.e, tlag, tau, fact,
dw = dw, method = method) CV.lo, CV.hi,
```

```
# 10,000 simulations, power model: C ~ a * t^b
# delta = 0.2 (acceptance limits {-delta, +delta})
# exponent of model = +0.25554
# last model’s pre-dose concentration = 15.78
# CV = 10.0%, median of exponents = +0.25334
# 97.8% positive exponents
# 2.2% negative exponents
# significant exponents: 11.55% (95% CI does not contain zero)
# failed exponents : 98.88% (95% CI not within acceptance limits)
# autocorrelation : 12.2%
# CV = 40.0%, median of exponents = +0.25455
# 69.9% positive exponents
# 30.1% negative exponents
# significant exponents: 5.23% (95% CI does not contain zero)
# failed exponents : 99.59% (95% CI not within acceptance limits)
# autocorrelation : 10.8%
```

Do we gain something? Not really.

As above the subset of 24 subjects.

We would arrive at the same conclusion than with the linear model.
When we run both models we see that in some cases the power model fits
better (based on an ANOVA or
the AIC). However, we
see also that some fits are *convex* and that is from a
theoretical perspective simply wrong. That’s a dead end.

So why are we not fitting the underlying function \[C=C_\textrm{ss}\big{(}1-\exp(-\lambda \times t)\big{)} \tag{5}\] of the build-up directly? There are two problems, a technical and a theoretical one.

We cannot linearize \(\small{(5)}\) like we did with the
power model \(\small{(3)}\). Hence, we require
nonlinear modeling as provided in R’s function
`nls()`

, which requires suitable start values. In many cases
\(\small{\max (C) \Rightarrow
C_\textrm{ss}}\) and \(\small{\log_{e}2\,/\,t_{1/2}\Rightarrow\lambda}\)
work. Alternatively we can set up the model in Phoenix WinNonlin.
However, in most cases calculating a
CI fails.

Even if we obtain a CI,
*which* parameter should we test and against *what*?

The true steady state is unknown and therefore, testing \(\small{\widehat{C_\textrm{ss}}}\) is not an
option. We face a similar problem with testing \(\small{\widehat{\lambda}}\). We have only
the half life used in study planning and no clue what the subjects’ true
predominant rate constants are.

“NCA and statistics associated with either C_{trough}or dosing interval PK parameters have limited value for SS characterization.

C_{trough}is not very sensitive to characterize steady state (SS), especially for compounds that attained steady state slowly.

Testing for a significant difference is as flawed as a similar
approach in the early 1980s in bioequivalence was.

A large difference with high variability passes and a small difference
with low variability fails:

Assessing the CI would only
make sense if there would be *a priori* defined acceptance
limits. Such limits were never defined. Even an extremely liberal \(\small{\Delta\,0.5}\) does not work because
the pre-dose concentrations are too variable. In the example above 83.73% would fail with low
variability and 97.46% with high variability.

As we have seen in the simulations, we can get caught in the trap of false positives. Consequently we would exclude subjects and loose power. On the other hand, esp. with high variability, we would keep subjects which are not in steady state.

Too few doses (say, based on five times the half life) might not be a
good idea. With low variability we will be punished by slopes
significantly different from zero and even with high variability the
plots will be telling.

This exercise is not just about making assessors happy. Based on the
superposition principle^{20} of
PK we __must__ compare
bioavailabilities __only__ after a single dose __or__ in
(pseudo-)steady state – not in the saturation phase between. Such a
comparison would be meaningless.

Recall the __legal__ definition of bioequivalence in the
U.S.A.:

“Two drug products will be considered bioequivalent drug products if they are pharmaceutical equivalents or pharmaceutical alternatives whose rate and extent of absorption do not show a significant difference when administered at the same molar dose of the active moiety under similar experimental conditions, either single dose or multiple dose.

Does that require testing for a __statistically__ significant
difference? Of course, not. It is meant in the simple language lawyers
can comprehend.

“significant:

Large enough to be noticed or have an effect, having or likely to have influence or effect.

Similarly here when it comes to ‘demonstration’.

“demonstration:

An event that proves a fact.

Recall that this was consistent with the EMA’s previous point of view

»*a* […] *statistical method is not considered necessary*
[…]. *Descriptive data is sufficient.*«

If you insist in using a statistical approach, estimate the
correlation between pre-dose concentrations by a mixed-effects model
with subject as a random effect.^{23} Good luck!

“[…] all models are approximations.

Essentially, all models are wrong, but some are useful.

**I hold that for ‘demonstrating achievement of steady state’
no statistical model is useful at all…**

The requirement of multiple dose studies is highly controversial. At the time being the EMA – apart from Australia – is the only jurisdiction requiring them. They were never necessary for the FDA, and Health Canada required them only for drugs with accumulation (\(\small{AUC_{0-\tau}>80\%\,AUC_{0-\infty}}\)) until 2012.

Extensive simulations^{25} showed that comparing the concentration
at the end of the intended dosing interval (\(\small{C_\tau}\)) – additionally to \(\small{AUC_{0-\text{t}}}\) and \(\small{C_\textrm{max}}\) – could reliably
predict the outcome after multiple doses.

The approach comes with a price: Since \(\small{C_\tau}\) is a one-point metric and hence, likely the one with the largest variability of the entire profile, the sample size has to be increased in order to preserve power. Nevertheless, the perspective to waive the multiple dose study is appealing.

It must be mentioned that the paper won the AAPS’ »Clinical Pharmacology and Translational Research«
*Outstanding Manuscript Award in Modeling and Simulation*

in 2012…

The proposal was challenged by a retrospective analysis^{26} of studies submitted
to the Spanish agency. The authors concluded that \(\small{C_\tau}\) after a single dose is a
poor predictor of \(\small{C_\textrm{min,ss}}\) after multiple
doses.

“[…] in […] six cases […] the multiple dose study was the only [sic] design able to detect the differences and, therefore, it was essential when comparing the in vivo performance of prolonged release products.

Regarding the predictive value of \(\small{C_\tau}\), one case […] shows that it is predictive of the bioequivalence failure of \(\small{C_\textrm{min,ss}}\), but in the other ﬁve cases, the results are not predictive or as sensitive as \(\small{C_\textrm{max,ss}}\) or \(\small{C_\textrm{min,ss}}\).

Regrettably, only failed studies were reported and therefore, it is
not possible to assess the FPR.
I’m not a friend of *post hoc*
power but evidently none [*sic*] of the studies was
sufficiently powered (median 11.84%, quartiles 3.25 – 13.35%) to
demonstrate BE for this metric. That
is not surprising, since assessing \(\small{C_\tau}\) was not a regulatory
requirement at submission. In short, the results do not refute but
rather *support* the simulation study: All cases which failed
after a single dose failed after multiple doses as well. Not only the
data base is obscure but – more importantly – the authors’ conclusions
contradict their findings. That’s bad in an age when many people don’t
read publications beyond its abstract. Alas, the paper is cited as
‘evidence’ to support the need of multiple dose studies.

The paper was defended^{27} and
IMHO, understandably,
challenged.^{28} ^{29} ^{30}

“Science should always be the basis of regulatory requirements.

Given the limited data on new PK
metrics, it was proposed at the
»EUFEPS
Open Discussion Forum«^{22} that regulations
should not be *made up out of thin air* but based on
*science*.

- BE should be assessed only by conventional PK metrics according to the (then) current guideline.
- Applicants should analyze studies with the suggested new
PK metrics in an
*exploratory*manner and submit results to agencies. - After a limited time frame (
*e.g.*, two years) the data could be assessed for their sensitivity and included in the new guideline if deemed necessary.^{32}

We all know how it ended. Although in April 2014^{33} it was promised that
»*the overview of comments will be published within the next two
weeks*«, we had to wait for 5½ years.^{34} ^{35} As usual, most
suggestions were ignored.

The Brazilian competent authority ANVISA
required multiple dose studies as well. The message was heard and hence,
the agency retrospectively assessed 117 studies submitted since 2008 for
early and late partial \(\small{AUC}\)
(\(\small{\textrm{p}AUC_{0-\tau/2}}\),
\(\small{\textrm{p}AUC_{\tau/2-\tau}}\)).^{36} 79.2%
of studies which passed conventional PK metrics, passed the partial
\(\small{AUC\textrm{s}}\) as well. In
order to pass, in general the sample size has not to be increased.
Therefore, the requirement was lifted and only \(\small{C_\text{max}}\), \(\small{AUC_{0-\text{t}}}\), \(\small{\textrm{p}AUC_{0-\tau/2}}\), \(\small{\textrm{p}AUC_{\tau/2-\tau}}\) after
a single dose are required.^{37} ^{38}

Let us hope that – in the spirit of global harmonization – sooner or later the EMA (and its ‘disciples’ in Australia) will follow all other agencies.

“In advanced engineering, you expected failure; you learned as much from failures as from successes – indeed if you never suffered a failure you probably weren’t pushing the envelope ambitiously enough.

Helmut Schütz 2022

`R`

GPL 3.0,
`klippy`

MIT,
`pandoc`

GPL 2.0.

1^{st} version January 16, 2022. Rendered September 12, 2022
12:29 CEST by rmarkdown via pandoc in 0.22 seconds.

Light RJ, Singer JD, Willett JB.

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*Guidance for Industry. Conduct and Analysis of Bioavailability and Bioequivalence Studies – Part B: Oral Modified Release Formulations.*Ottawa. 1996. Internet Archive.↩︎EMEA.

*Overview of Comments received on Draft Guideline on the Investigation of Bioequivalence.*London. 20 January 2010. Online.↩︎Health Canada.

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*Draft Guidance for Industry. Bioequivalence Studies With Pharmacokinetic Endpoints for Drugs Submitted Under an ANDA.*Silver Spring. August 2021. Download.↩︎Fang L.

*PK and Statistical Considerations for Steady State BE Studies – FDA Perspective*. Presentation at SAAM*now*Spring 2019 Workshop. Rockville. April 4–5, 2019.↩︎EMA, CHMP.

*Guideline on the pharmacokinetic and clinical evaluation of modified release dosage forms.*London. 20 November 2014. Online.↩︎TGA.

*International scientific guidelines adopted in Australia. Guideline on the pharmacokinetic and clinical evaluation of modified release dosage forms with TGA annotations.*1 Jun 2015. Online.↩︎Weimann HJ.

*Drug concentrations and directly derived parameters.*In: Cawello W (editor).*Parameters for Compartment-Free Pharmacokinetics.*Aachen: Shaker-Verlag; 2003. p. 32.↩︎I have to confess that I have recommended it in the past.↩︎

Wolfsegger MJ, Bauer A, Labes D, Schütz H, Vonk R, Lang B, Lehr S, Jaki TF, Engl W, Hale MD.

*Assessing goodness-of-ﬁt for evaluation of dose-proportionality.*Pharm Stat. 2021; 20(2): 272–81. doi:10.1002/pst.2074.↩︎Nicolas X, Sultan E, Ollier C, Rauch C, Fabre D.

*Estimation by different techniques of steady-state achievement and accumulation for a compound with bi-phasic disposition and long terminal half-life.*PAGE meeting. Montpellier. July 7–13, 2013. Online.↩︎Dost FH.

*Der Blutspiegel. Konzentrationsabläufe in der Kreislaufflüssigkeit.*Leipzig: VEB Thieme; 1953. p. 244. [German]↩︎CFR. Title 21. Volume 5. Chapter I. §320.23 (b)(1)

*Basis for measuring in vivo bioavailability or demonstrating bioequivalence.*Dec. 19, 2002. Online.↩︎EUFEPS.

*Open Discussion Forum on the Revised European Guideline on Pharmacokinetic and Clinical Evaluation of Modified Release Dosage Forms.*Bonn. 17–18 June, 2013.↩︎Maganti L, Panebianco DL, Maes AL.

*Evaluation of Methods for Estimating Time to Steady State with Examples from Phase 1 Studies.*AAPS J. 2008; 10(1): 141–7. doi:10.1208%2Fs12248-008-9014-y. Free Full Text.↩︎Box GEP, Draper NR.

*Empirical Model-Building and Response Surfaces.*New York: Wiley; 1987. p. 424.↩︎Paixão P, Gouveia LF, Morais JAG.

*An alternative single dose parameter to avoid the need for steady-state studies on oral extended-release drug products.*Eur J Pharmaceut Biopharmaceut. 2012; 80(2): 410–7. doi:10.1016/j.ejpb.2011.11.001.↩︎García-Arieta A, Morales-Alcelay S, Herranz M, de la Torre-Alvarado JM, Blázquez-Pérez A, Suárez-Gea ML, Alvarez C.

*Investigation on the need of multiple dose bioequivalence studies for prolonged-release generic products.*Int J Pharm. 2012; 423(2): 321–5. doi:10.1016/j.ijpharm.2011.11.022.↩︎García‐Arieta A.

*Justification of the current regulatory approach by EMA prohibiting the extrapolation of single dose BE to steady state in many cases.*GBHI, 3^{rd}International Workshop. Amsterdam. April 12, 2018.↩︎Schütz H.

*Primary and secondary PK metrics for evaluation of steady state studies,**C*_{min}*vs.**C*_{τ},*relevance of**C*_{min}/*C*_{τ}*or fluctuation for bioequivalence assessment.*GBHI, 3^{rd}International Workshop. Amsterdam. April 12, 2018.↩︎Silva N.

*Suggestions for Further Harmonization of Remaining Open Issues.*GBHI, 4^{th}International Workshop. Bethesda. December 12, 2019.↩︎Cook J.

*Scientific Arguments in Favor and Against the Requirement to Perform Multiple Dose Studies for Long Acting Injectables and Implants.*GBHI, 4^{th}International Workshop. Bethesda. December 12, 2019.↩︎Röhmel J.

*Recent Methodological Advances Contributing to Clinical Trials and Regulatory Statistics.*30^{th}Annual Conference of the International Society for Clinical Biostatistics. Prague. August 25, 2009.↩︎That’s similar to what the FDA did when concerns about the ‘Subject-by-Formulation’ interaction emerged. Temporarily studies had to be performed in a replicated design. After assessing the results, the FDA concluded that the S × F interaction is not relevant and the temporary requirement was lifted.↩︎

EMA/EGA.

*Joint workshop on the impact of the revised EMA Guideline on the Pharmacokinetic and Clinical Evaluation of Modified Release Dosage Forms.*London. 30 April 2014.↩︎EMA, PKWP.

*Overview of comments received on ‘Guideline on the pharmacokinetic and clinical evaluation of modified release dosage forms’.*EMA/CHMP/267430/2019. London. 17 October 2019. Online.↩︎Regrettably it does not contain a column ‘Outcome’, which served well in other documents in understanding the EMA’s point of view.↩︎

Costa Soares KE, Chiann C, Storpitis S.

*Assessment of the impact of partial area under the curve in a bioavailability/bioequivalence study on generic prolonged-release formulations.*Eur J Pharm Sci. 2022; 171: 106127. doi:10.1016/j.ejps.2022.106127.↩︎Costa Soares KE (ANVISA/CETER). Presentation at:

*Multiple dose studies for generic and similar drugs of modified release.*São Paulo. April 29, 2021. URL.↩︎ANVISA. Resolução - RDC Nº 742.

*Dispõe sobre os critérios para a condução de estudos de biodisponibilidade relativa/bioequivalência (BD/BE) e estudos farmacocinéticos.*Brasilia. August 10, 2022. Effective July 3, 2023. Online. [Portuguese]↩︎Baxter S.

*Transcendent.*London: Gollanz; 2005. Chapter 36.↩︎